Statistical notes

Miscellaneous statistical stuff (statistics, decision theory)
created: 17 July 2014; modified: 02 Apr 2017; status: in progress; confidence: possible;

Critiques

• criticism of teeth-removal experiment in rats http://lesswrong.com/r/discussion/lw/kfb/open_thread_30_june_2014_6_july_2014/b1u3
• criticism of small Noopept self-experiment http://www.bluelight.org/vb/threads/689936-My-Paper-quot-Noopept-amp-The-Placebo-Effect-quot?p=11910708&viewfull=1#post11910708
• why Soylent is not a good idea http://lesswrong.com/lw/hht/link_soylent_crowdfunding/90y7
• misinterpretation of fluoridation meta-analysis and ignorance of VoI http://theness.com/neurologicablog/index.php/anti-fluoride-propaganda-as-news/#comment-76400
• http://lesswrong.com/lw/1lt/case_study_melatonin/8mgf
• Fulltext: https://dl.dropboxusercontent.com/u/280585369/2014-dubal.pdf is this possible? http://nextbigfuture.com/2014/05/kl-vs-gene-makes-up-six-iq-points-of.html#comment-1376748788 http://www.reddit.com/r/Nootropics/comments/25233r/boost_your_iq_by_6_points/chddd7f
• tACS causes lucid dreaming: http://www.reddit.com/r/LucidDreaming/comments/27y7n6/no_brain_stimulation_will_not_get_you_lucid/ck6isgo
• slave whippings vs cotton production http://lesswrong.com/r/discussion/lw/kwc/open_thread_sept_17_2014/bajv
• whether a study on mental illness & violence shows schizophrenics are not more likely to murder but rather be murdered: http://www.reddit.com/r/psychology/comments/2fwjs8/people_with_mental_illness_are_more_likely_to_be/ckdq50k / http://www.nationalelfservice.net/publication-types/observational-study/people-with-mental-illness-are-more-likely-to-be-victims-of-homicide-than-perpetrators-of-homicide/#comment-95507 (see also http://slatestarscratchpad.tumblr.com/post/120950150581/psycholar-giraffepoliceforce-museicetc )
• Fortune analysis of higher female CEO returns http://lesswrong.com/r/discussion/lw/l3b/contrarian_lw_views_and_their_economic/bftw
• failed attempt at estimating P(causation|correlation) https://plus.google.com/103530621949492999968/posts/UzMMmPgyyaV
• vitamin D/caffeine claim based on weak in vitro claims, inconsistent with more relevant in vivo results: https://plus.google.com/u/0/103530621949492999968/posts/AUg3udezXMS [already included in Nootropics.page]
• claimed fall in digit span backwards minuscule and non-statistically-significant, no evidence of heterogeneity beyond variability due to sample size http://drjamesthompson.blogspot.com/2015/04/digit-span-bombshell.html?showComment=1428096775425#c4097303932864318518
• Claimed randomized experiment of whether sushi tastes worse after freezing is not actually a randomized experiment https://www.reddit.com/r/science/comments/324xmf/randomized_doubleblind_study_shows_the_quality_of/cq8dmsb
• aerobic vs weightlifting exercise, multiple problems but primarily p-hacking, difference-in-statistical-significance-is-not-a-significant-difference, and controlling for intermediate variable: https://plus.google.com/103530621949492999968/posts/aeZqB8JWUiQ

• another controlling for intermediate variable: https://twitter.com/gwern/status/538763784048095232
• sexual openness result undermined by ceiling effect http://mindhacks.com/2015/04/28/when-society-isnt-judging-womens-sex-drive-rivals-mens/#comment-362749
• music study claiming WM interaction: possible ceiling effect? see FB PM
• attempt to measure effect of Nazi anti-schizophrenia eugenics program failed to use breeder’s equation to estimate possible size of effect, which is too small to detect with available data and hence attempt is foredoomed: https://www.reddit.com/r/eugenics/comments/3hqdll/between_73_and_100_of_all_individuals_with/cul2nzw
• claim high IQ types almost 100% failure rates due to inappropriate model assumption of normal distribution + narrow standard deviation: http://polymatharchives.blogspot.com/2015/01/the-inappropriately-excluded.html?showComment=1441741719623#c1407914596750199739
• implausible claims about success rate of facial recognition applied to St Petersburg population: https://news.ycombinator.com/item?id=11491264 (see also Facial recognition systems stumble when confronted with million-face database)
• human Toxoplasma gondii study is not well-powered as authors claim due to incorrect power analysis, and results are evidence for harm: http://blogs.discovermagazine.com/neuroskeptic/2016/02/20/myth-mind-altering-parasite-toxoplasma-gondii/#comment-2755778490 ; https://www.reddit.com/r/slatestarcodex/comments/5vjrmo/toxoplasma_doesnt_cause_adolescent_psychosis/de2x4kh/
• attempt at attributing Bitcoin price increases to technology improvements: https://www.reddit.com/r/Bitcoin/comments/5tbt8f/buzz_factor_or_innovation_potential_what_explains/ddlmzrz/

Someone Should Do Something: Wishlist of Miscellaneous Project Ideas

• Erowid: data-mine the trip reports to create clusters or a state-space of drug effects/results and using the clusters & common terms, create a general inventory of descriptions; add this to the trip report form so Erowid users can provide some more structured information about their experience.

• Dark net markets: use a longitudinal crawl of DNM sellers to estimate survival curves, outstanding escrow + orders, and listed product prices / type / language to try to predict exit scams.
• Dark net markets: parse hyperlinks in all forums & DNMs and outlinked domains like Reddit’s /r/darknetmarkets in order to create a proper network topology of Tor hidden services - existing .onion crawls tend to find Tor hidden services are a very fragmented and unconnected network despite the Hidden Wikis, but this may reflect the instability of hidden services and the use of clearnet communities & indexes (like Reddit or gwern.net)

• read through the Extropy and SL4 mailing list archives: have any important ideas been forgotten or dropped? Are there any meta lessons about forecasting or thinking about existential risk to be gleaned?

• an efficient hashmap library for R, perhaps binding to a C/C++ library; existing R hashmaps are based on environments and not very fast

• key for enabling many statistical and especially reinforcement learning algorithms like Monte Carlo tree search (MCTS)
• Markov chain/char-RNN bot for Twitter trained on just English proverbs, idioms, and expressions
• user-friendly char-RNN implementation just for classifying text, taking in CSV data of text/category
• hyperparameter optimization for algorithms in problems without available loss functions but human-judgeable quality, using a human for making choices in a paired or forced-choice comparison, then using a Bradley-Terry or latent variable model to infer rankings of hyperparameter settings and optimizing based on the latent scores

• a VR application for viewing stereoscopic images & video and for 3D environments with extremely large parallax such as for viewing clouds with true depth perception (discussion)
• properly transcribe & annotate Douglas Hofstader’s Le Ton Beau de Marot, one of his best but also most obscure books

• provide polygenic scores as a service, a website/API where one can upload a SNP data file like the 23andMe export and get back PGSes for everything in LD Hub, and utility weights
• track down & upload missing papers in http://arthurjensen.net/?page_id=9
• expand/rewrite Wikipedia’s heritability of intelligence - grossly outdated, almost totally omitting all the GCTAs and GWASes that have definitively settled the answer in the strongly affirmative

Estimating censored test scores

An acquaintance asks the following question: he is applying for a university course which requires a certain minimum score on a test for admittance, and wonders about his chances and a possible trend of increasing minimum scores over time. (He hasn’t received his test results yet.) The university doesn’t provide a distribution of admittee scores, but it does provide the minimum scores for 2005-2013, unless all applicants were admitted because they all scored above an unknown cutoff - in which case it provides no minimum score. This leads to the dataset:

2005,NA
2006,410
2007,NA
2008,NA
2009,398
2010,407
2011,417
2012,NA
2013,NA

A quick eyeball tells us that we can’t conclude much: only 4 actual datapoints, with 5 hidden from us. We can’t hope to conclude anything about time trends, other than there doesn’t seem to be much of one: the last score, 417, is not much higher than 410, and the last two scores are low enough to be hidden. We might be able to estimate a mean, though.

We can’t simply average the 4 scores and conclude the mean minimum is 410 because of those NAs: a number of scores have been censored because they were too low, and while we don’t know what they were, we do know they were <398 (the smallest score) and so a bunch of <398s will pull down the uncensored mean of 410.

On approach is to treat it as a Tobit model and estimate using something like the censReg library (overview).

But if we try a quick call to censReg, we are confounded: a Tobit model expects you to provide the cutoff below which the observations were censored, but that is something we don’t know. All we know is that it must be below 398, we weren’t told it was exactly 395, 394, etc. Fortunately, this is a solved problem. For example: The Tobit model with a non-zero threshold, Carson & Sun 2007 tells us:

In this paper, we consider estimating the unknown censoring threshold by the minimum of the uncensored $y_i$’s. We show that the estimator $γ'$ of $γ$ is superconsistent and asymptotically exponentially distributed. Carson (1988, 1989) also suggests estimating the unknown censoring threshold by the minimum of the uncensored $y_i$’s. In a recent paper, Zuehlke (2003) rediscovers these unpublished results and demonstrates via simulations that the asymptotic distribution of the maximum likelihood estimator does not seem to be affected by the estimation of the censoring threshold.

That seems to be almost too simple and easy, but it makes sense and reminds me a little of the German tank problem: the minimum might not be that accurate a guess (it’s unlikely you just happened to draw a sample right on the censoring threshold) and it definitely can’t be wrong in the sense of being too low. (A Bayesian method might be able to do better with a prior like a exponential.)

With that settled, the analysis is straightforward: load the data, figure out the minimum score, set the NAs to 0, regress, and extract the model estimates for each year:

scores <- data.frame(Year=2005:2013,
MinimumScore=c(NA,410,NA,NA,398,407,417,NA,NA));
censorThreshold <- min(scores$MinimumScore, na.rm=T) scores[is.na(scores)] <- 0 library(censReg) # 'censorThreshold-1' because censReg seems to treat threshold as < and not <= summary(censReg(MinimumScore ~ Year, left=censorThreshold-1, data=scores)) # Warning message: # In censReg(MinimumScore ~ Year, left = censorThreshold - 1, data = scores) : # at least one value of the endogenous variable is smaller than the left limit # # Call: # censReg(formula = MinimumScore ~ Year, left = censorThreshold - # 1, data = scores) # # Observations: # Total Left-censored Uncensored Right-censored # 9 5 4 0 # # Coefficients: # Estimate Std. error t value Pr(> t) # (Intercept) -139.9711 Inf 0 1 # Year 0.2666 Inf 0 1 # logSigma 2.6020 Inf 0 1 # # Newton-Raphson maximisation, 37 iterations # Return code 1: gradient close to zero # Log-likelihood: -19.35 on 3 Df -139.9711 + (0.2666 * scores$Year)
# [1] 394.6 394.8 395.1 395.4 395.6 395.9 396.2 396.4 396.7

With so little data the results aren’t very reliable, but there is one observation we can make.

The fact that half the dataset is censored tells us that the uncensored mean may be a huge overestimate (since we’re only looking at the top half of the underlying data), and indeed it is. The original mean of the uncensored scores was 410; however, the estimate including the censored data is much lower, 397 (13 less)!

This demonstrates the danger of ignoring systematic biases in your data.

So, trying to calculate a mean or time effect is not helpful. What might be better is to instead exploit the censoring directly: if the censoring happened because everyone got in, then if you showed up in a censored year, you have 100% chance of getting in; while in a non-censored year you have an unknown but <100% chance of getting in; so the probability of a censored year sets a lower bound on one’s chances, and this is easy to calculate as a simple binomial problem - 5 out of 9 years were censored years, so:

binom.test(c(5,4))
#
#   Exact binomial test
#
# data:  c(5, 4)
# number of successes = 5, number of trials = 9, p-value = 1
# alternative hypothesis: true probability of success is not equal to 0.5
# 95% confidence interval:
#  0.212 0.863
# sample estimates:
# probability of success
#                 0.5556

So we can tell him that he may have a >55% chance of getting in.

The Traveling Gerontologist problem

A quick probability exercise: Wikipedia mentions Finland has 566 centenarians as of 2010.

That’s few enough you could imagine visiting them all to research them and their longevity, in a sort of traveling salesman problem but with gerontologists instead. Except, because of the exponential increase in mortality, centenarians have high annual mortality rates; it depends on the exact age but you could call it >30% (eg Finnish 99yos in 2012 had a death toll of 326.54/1000). So you might well try to visit a centenarian and discover they’d died before you got there.

How bad a risk is this? Well, if the risk per year is 30%, then one has a 70% chance of surviving a year. To survive a year, you must survive all 365 days; by the multiplication rule, the risk is $x$ where $0.7 = x \cdot x \cdot x \cdot ... * x \text{[365.25 times]}$ or $0.7 = x^{365.25}$; solving, $x = 0.999024$.

It takes time to visit a centenarian - it wouldn’t do to be abrupt and see them for only a few minutes, you ought to listen to their stories, and you need to get to a hotel or airport, so let’s assume you visit 1 centenarian per day.

If you visit centenarian A on day 1, and you want to visit centenarian B on day 2, then you can count on a 99.9% chance B is still alive. So far so good. And if you wanted to visit 566 centenarians (let’s imagine you have a regularly-updated master list of centenarians from the Finnish population registry), then you only have to beat the odds 566 times in a row, which is not that hard: $0.999024^{566} = 0.5754023437943274$.

But that’s coldblooded of you to objectify those Finnish centenarians! Any centenarian will do, I don’t care. What if you picked the current set of 566 centenarians and wanted to visit just them, specifically - with no new centenarians introduced to the list to replace any dead ones.

That’s a little more complicated. When you visit the first centenarian, it’s the same probability: 0.999024. When you visit the second centenarian the odds change since now she (and it’s more often she than he, since remember the exponential and males having shorter mean lifetimes) has to survive 2 days, so it’s $0.999024 \cdot 0.999024$ or $0.999024^2$; for the third, it’s $0.999024^3$, and so on to #566 who has been patiently waiting and trying to survive a risk of $0.999024^566$, and then you need to multiply to get your odds of beating every single risk of death and the centenarian not leaving for a more permanent rendezvous: $0.999024 \cdot 0.999024^2 \cdot 0.999024^3 \cdot ... \cdot 0.999024^{566}$, which would be $\prod_{n=1}^{566} 0.999024^n$, or in Haskell:

product (map (\x -> 0.999024**x) [1..566])
~> 8.952743340164081e-69

(A little surprisingly, Wolfram Alpha can solve the TeX expression too.)

Given the use of floating point in that function (567 floating point exponentiations followed by as many multiplications) and the horror stories about floating point, one might worry the answer is wrong & the real probability is much larger. We can retry with an implementation of computable reals, CReal, which can be very slow but should give more precise answers:

:module + Data.Number.CReal
showCReal 100 (product (map (\x -> 0.999024**x) [1..566]))
~> 0.0000000000000000000000000000000000000000000000000000000000000000000089527433401308585720915431195262

Looks good - agrees with the floating point version up to the 11th digit:

8.9527433401 64081e-69
8.9527433401 308585720915431195262

We can also check by rewriting the product equation to avoid all the exponentiation and multiplication (which might cause issues) in favor of a single exponential:

1. $p^1 * p^2 * ... p^n$ (as before)
2. = $p^{1+2+...+n}$ (since $(x^m) * (x^n) = x^(m + n)$)
3. = $p^{\frac{n \cdot (1 + n)}{2}}$ (by arithmetic progression/Gauss’s famous classroom trick since $\sum_1^n = n \cdot \frac{a_1 + a_n}{2}$)
4. = $0.999024^{\frac{566 \cdot (1 + 566)}{2}}$ (start substituting in specific values)
5. = $0.999024^{\frac{320922}{2}}$
6. = $0.999024^{160461}$

So:

0.999024^160461
~> 8.95274334014924e-69

Or to go back to the longer version:

0.999024**((566*(1 + 566)) / 2)
~> 8.952743340164096e-69

Also close. All probabilities of success are minute.

How fast would you have to be if you wanted to at least try to accomplish the tour with, say, a 50-50 chance?

Well, that’s easy: you can consider the probability of all of them surviving one day and as we saw earlier, that’s $0.999024^{566} = 0.58$, and two days would be $(0.999024 ^ {566}) ^ 2 = 0.33$ So you can only take a little over a day before you’ve probabilistically lost & one of them has died; if you hit all 566 centenarians in 24 hours, that’s ~24 centenarians per hour or ~2 minutes to chat with each one and travel to the next. If you’re trying to collect DNA samples, better hope they’re all awake and able to give consent!

So safe to say, you will probably not be able to manage the Traveling Gerontologist’s tour.

Bayes nets

Daily weight data graph

As the datasets I’m interested in grow in number of variables, it becomes harder to justify doing analysis by simply writing down a simple linear model with a single dependent variable and throwing in the independent variables and maybe a few transformations chosen by hand. I can instead write down some simultaneous-equations/structural-equation-models, but while it’s usually obvious what to do for k<4 and if it’s not I can compare the possible variants, 4 variables is questionable what the right SEM is, and >5, it’s hopeless. Factor analysis to extract some latent variables is a possibility, but the more general solution here seems to be probabilistic graphical models such as Bayesian networks.

I thought I’d try out some Bayes net inference on some of my datasets. In this case, I have ~150 daily measurements from my Omron body composition scale, measuring total weight, body fat percentage, and some other things (see an Omron manual):

1. Total weight
2. BMI
3. Body fat percentage
4. Muscle percentage
5. Resting metabolism in calories
6. Body age
7. Visceral fat index

The 7 variables are interrelated, so this is definitely a case where a simple lm is not going to do the trick. It’s also not 100% clear how to set up a SEM; some definitions are obvious (the much-criticized BMI is going to be determined solely by total weight, muscle and fat percentage might be inversely related) but others are not (how does visceral fat relate to body fat?). And it’s not a hopelessly small amount of data.

The Bayes net R library I’m trying out is bnlearn (paper).

library(bnlearn)
# https://www.dropbox.com/s/4nsrszm85m47272/2015-03-22-gwern-weight.csv
weight$Date <- NULL; weight$Weight.scale <- NULL
# remove missing data
weightC <- na.omit(weight)
# bnlearn can't handle integers, oddly enough
weightC <- as.data.frame(sapply(weightC, as.numeric))
summary(weightC)
#   Weight.Omron        Weight.BMI        Weight.body.fat    Weight.muscle
#  Min.   :193.0000   Min.   : 26.90000   Min.   :27.00000   Min.   :32.60000
#  1st Qu.:195.2000   1st Qu.: 27.20000   1st Qu.:28.40000   1st Qu.:34.20000
#  Median :196.4000   Median : 27.40000   Median :28.70000   Median :34.50000
#  Mean   :196.4931   Mean   : 28.95409   Mean   :28.70314   Mean   :34.47296
#  3rd Qu.:197.8000   3rd Qu.: 27.60000   3rd Qu.:29.10000   3rd Qu.:34.70000
#  Max.   :200.6000   Max.   : 28.00000   Max.   :31.70000   Max.   :35.50000
#  Weight.resting.metabolism Weight.body.age    Weight.visceral.fat
#  Min.   :1857.000          Min.   :52.00000   Min.   : 9.000000
#  1st Qu.:1877.000          1st Qu.:53.00000   1st Qu.:10.000000
#  Median :1885.000          Median :53.00000   Median :10.000000
#  Mean   :1885.138          Mean   :53.32704   Mean   : 9.949686
#  3rd Qu.:1893.000          3rd Qu.:54.00000   3rd Qu.:10.000000
#  Max.   :1914.000          Max.   :56.00000   Max.   :11.000000
cor(weightC)
#                             Weight.Omron     Weight.BMI Weight.body.fat  Weight.muscle
# Weight.Omron               1.00000000000  0.98858376919    0.1610643221 -0.06976934825
# Weight.BMI                 0.98858376919  1.00000000000    0.1521872557 -0.06231142104
# Weight.body.fat            0.16106432213  0.15218725566    1.0000000000 -0.98704369855
# Weight.muscle             -0.06976934825 -0.06231142104   -0.9870436985  1.00000000000
# Weight.resting.metabolism  0.96693236051  0.95959140245   -0.0665001241  0.15621294274
# Weight.body.age            0.82581939626  0.81286141659    0.5500409365 -0.47408608681
# Weight.visceral.fat        0.41542744168  0.43260100665    0.2798756916 -0.25076619829
#                           Weight.resting.metabolism Weight.body.age Weight.visceral.fat
# Weight.Omron                           0.9669323605    0.8258193963        0.4154274417
# Weight.BMI                             0.9595914024    0.8128614166        0.4326010067
# Weight.body.fat                       -0.0665001241    0.5500409365        0.2798756916
# Weight.muscle                          0.1562129427   -0.4740860868       -0.2507661983
# Weight.resting.metabolism              1.0000000000    0.7008354776        0.3557229425
# Weight.body.age                        0.7008354776    1.0000000000        0.4840752389
# Weight.visceral.fat                    0.3557229425    0.4840752389        1.0000000000

## create alternate dataset expressing the two percentage variables as pounds, since this might fit better
weightC2 <- weightC
weightC2$Weight.body.fat <- weightC2$Weight.Omron * (weightC2$Weight.body.fat / 100) weightC2$Weight.muscle   <- weightC2$Weight.Omron * (weightC2$Weight.muscle / 100)

Begin analysis:

pdap <- hc(weightC)
pdapc2 <- hc(weightC2)
## bigger is better:
score(pdap, weightC)
# [1] -224.2563072
score(pdapc2, weightC2)
# [1] -439.7811072
## stick with the original, then
pdap
#   Bayesian network learned via Score-based methods
#
#   model:
#    [Weight.Omron][Weight.body.fat][Weight.BMI|Weight.Omron]
#    [Weight.resting.metabolism|Weight.Omron:Weight.body.fat]
#    [Weight.body.age|Weight.Omron:Weight.body.fat]
#    [Weight.muscle|Weight.body.fat:Weight.resting.metabolism][Weight.visceral.fat|Weight.body.age]
#   nodes:                                 7
#   arcs:                                  8
#     undirected arcs:                     0
#     directed arcs:                       8
#   average markov blanket size:           2.57
#   average neighbourhood size:            2.29
#   average branching factor:              1.14
#
#   learning algorithm:                    Hill-Climbing
#   score:                                 BIC (Gauss.)
#   penalization coefficient:              2.534452101
#   tests used in the learning procedure:  69
#   optimized:                             TRUE
plot(pdap)
## https://i.imgur.com/nipmqta.png

This inferred graph is obviously wrong in several respects, violating prior knowledge about some of the relationships.

More specifically, my prior knowledge:

• Weight.Omron == total weight; should be influenced by Weight.body.fat (%), Weight.muscle (%), & Weight.visceral.fat
• Weight.visceral.fat: ordinal variable, <=9 = normal; 10-14 = high; 15+ = very high; from the Omron manual:

Visceral fat area (0 - approx. 300 cm , 1 inch=2.54 cm) distribution with 30 levels. NOTE: Visceral fat levels are relative and not absolute values.

• Weight.BMI: BMI is a simple function of total weight & height (specifically BMI = round(weight / height^2)), so it should be influenced only by Weight.Omron, and influence nothing else
• Weight.body.age: should be influenced by Weight.Omron, Weight.body.fat, and Weight.muscle, based on the description in the manual:

Body age is based on your resting metabolism. Body age is calculated by using your weight, body fat percentage and skeletal muscle percentage to produce a guide to whether your body age is above or below the average for your actual age.

• Weight.resting.metabolism: a function of the others, but I’m not sure which exactly; manual talks about what resting metabolism is generically and specifies it has the range 385 to 3999 kcal with 1 kcal increments; https://en.wikipedia.org/wiki/Basal_metabolic_rate suggests the Omron may be using one of several approximation equations based on age/sex/height/weight, but it might also be using lean body mass as well.

Unfortunately, bnlearn doesn’t seem to support any easy way of encoding the prior knowledge - for example, you can’t say no outgoing arrows from node X - so I iterate, adding bad arrows to the blacklist.

Which arrows violate prior knowledge?

• [Weight.visceral.fat|Weight.body.age] (read backwards, as Weight.body.age ~> Weight.visceral.fat)
• [Weight.muscle|Weight.resting.metabolism]

Retry, blacklisting those 2 arrows:

pdap2 <- hc(weightC, blacklist=data.frame(from=c("Weight.body.age", "Weight.resting.metabolism"), to=c("Weight.visceral.fat","Weight.muscle")))

New violations:

• [Weight.visceral.fat|Weight.BMI]
• [Weight.muscle|Weight.Omron]
pdap3 <- hc(weightC, blacklist=data.frame(from=c("Weight.body.age", "Weight.resting.metabolism", "Weight.BMI", "Weight.Omron"), to=c("Weight.visceral.fat","Weight.muscle", "Weight.visceral.fat", "Weight.muscle")))

New violations:

• [Weight.visceral.fat|Weight.Omron]
• [Weight.muscle|Weight.BMI]
pdap4 <- hc(weightC, blacklist=data.frame(from=c("Weight.body.age", "Weight.resting.metabolism", "Weight.BMI", "Weight.Omron", "Weight.Omron", "Weight.BMI"), to=c("Weight.visceral.fat","Weight.muscle", "Weight.visceral.fat", "Weight.muscle", "Weight.visceral.fat", "Weight.muscle")))

One violation:

• [Weight.muscle|Weight.body.age]
pdap5 <- hc(weightC, blacklist=data.frame(from=c("Weight.body.age", "Weight.resting.metabolism", "Weight.BMI", "Weight.Omron", "Weight.Omron", "Weight.BMI", "Weight.body.age"), to=c("Weight.visceral.fat","Weight.muscle", "Weight.visceral.fat", "Weight.muscle", "Weight.visceral.fat", "Weight.muscle", "Weight.muscle")))
#   Bayesian network learned via Score-based methods
#
#   model:
#    [Weight.body.fat][Weight.muscle|Weight.body.fat][Weight.visceral.fat|Weight.body.fat]
#    [Weight.Omron|Weight.visceral.fat][Weight.BMI|Weight.Omron]
#    [Weight.resting.metabolism|Weight.Omron:Weight.body.fat]
#    [Weight.body.age|Weight.Omron:Weight.body.fat]
#   nodes:                                 7
#   arcs:                                  8
#     undirected arcs:                     0
#     directed arcs:                       8
#   average markov blanket size:           2.57
#   average neighbourhood size:            2.29
#   average branching factor:              1.14
#
#   learning algorithm:                    Hill-Climbing
#   score:                                 BIC (Gauss.)
#   penalization coefficient:              2.534452101
#   tests used in the learning procedure:  62
#   optimized:                             TRUE
plot(pdap5)
## https://i.imgur.com/nxCfmYf.png

## implementing all the prior knowledge cost ~30:
score(pdap5, weightC)
# [1] -254.6061724

No violations, so let’s use the network and estimate the specific parameters:

fit <- bn.fit(pdap5, weightC); fit
#   Bayesian network parameters
#
#   Parameters of node Weight.Omron (Gaussian distribution)
#
# Conditional density: Weight.Omron | Weight.visceral.fat
# Coefficients:
#         (Intercept)  Weight.visceral.fat
#       169.181651376          2.744954128
# Standard deviation of the residuals: 1.486044472
#
#   Parameters of node Weight.BMI (Gaussian distribution)
#
# Conditional density: Weight.BMI | Weight.Omron
# Coefficients:
#   (Intercept)   Weight.Omron
# -0.3115772322   0.1411044216
# Standard deviation of the residuals: 0.03513413381
#
#   Parameters of node Weight.body.fat (Gaussian distribution)
#
# Conditional density: Weight.body.fat
# Coefficients:
# (Intercept)
# 28.70314465
# Standard deviation of the residuals: 0.644590085
#
#   Parameters of node Weight.muscle (Gaussian distribution)
#
# Conditional density: Weight.muscle | Weight.body.fat
# Coefficients:
#     (Intercept)  Weight.body.fat
#   52.1003347352    -0.6141270921
# Standard deviation of the residuals: 0.06455478599
#
#   Parameters of node Weight.resting.metabolism (Gaussian distribution)
#
# Conditional density: Weight.resting.metabolism | Weight.Omron + Weight.body.fat
# Coefficients:
#     (Intercept)     Weight.Omron  Weight.body.fat
#   666.910582196      6.767607964     -3.886694779
# Standard deviation of the residuals: 1.323176507
#
#   Parameters of node Weight.body.age (Gaussian distribution)
#
# Conditional density: Weight.body.age | Weight.Omron + Weight.body.fat
# Coefficients:
#     (Intercept)     Weight.Omron  Weight.body.fat
#  -32.2651379176     0.3603672788     0.5150134225
# Standard deviation of the residuals: 0.2914301529
#
#   Parameters of node Weight.visceral.fat (Gaussian distribution)
#
# Conditional density: Weight.visceral.fat | Weight.body.fat
# Coefficients:
#     (Intercept)  Weight.body.fat
#    6.8781100009     0.1070118125
# Standard deviation of the residuals: 0.2373649058
## residuals look fairly good, except for Weight.resting.metabolism, where there are some extreme residuals in what looks a bit like a sigmoid sort of pattern, suggesting nonlinearities in the Omron scale's formula?
bn.fit.qqplot(fit)
## https://i.imgur.com/mSallOv.png

We can double-check the estimates here by turning the Bayes net model into a SEM and seeing how the estimates compare, and also seeing if the p-values suggest we’ve found a good model:

library(lavaan)
Weight.model1 <- '
Weight.visceral.fat ~ Weight.body.fat
Weight.Omron ~ Weight.visceral.fat
Weight.BMI ~ Weight.Omron
Weight.body.age ~ Weight.Omron + Weight.body.fat
Weight.muscle ~ Weight.body.fat
Weight.resting.metabolism ~ Weight.Omron + Weight.body.fat
'
Weight.fit1 <- sem(model = Weight.model1,  data = weightC)
summary(Weight.fit1)
# lavaan (0.5-16) converged normally after 139 iterations
#
#   Number of observations                           159
#
#   Estimator                                         ML
#   Minimum Function Test Statistic               71.342
#   Degrees of freedom                                 7
#   P-value (Chi-square)                           0.000
#
# Parameter estimates:
#
#   Information                                 Expected
#   Standard Errors                             Standard
#
#                    Estimate  Std.err  Z-value  P(>|z|)
# Regressions:
#   Weight.visceral.fat ~
#     Weight.bdy.ft     0.107    0.029    3.676    0.000
#   Weight.Omron ~
#     Wght.vscrl.ft     2.745    0.477    5.759    0.000
#   Weight.BMI ~
#     Weight.Omron      0.141    0.002   82.862    0.000
#   Weight.body.age ~
#     Weight.Omron      0.357    0.014   25.162    0.000
#     Weight.bdy.ft     0.516    0.036   14.387    0.000
#   Weight.muscle ~
#     Weight.bdy.ft    -0.614    0.008  -77.591    0.000
#   Weight.resting.metabolism ~
#     Weight.Omron      6.730    0.064  104.631    0.000
#     Weight.bdy.ft    -3.860    0.162  -23.837    0.000
#
# Covariances:
#   Weight.BMI ~~
#     Weight.body.g    -0.000    0.001   -0.116    0.907
#     Weight.muscle    -0.000    0.000   -0.216    0.829
#     Wght.rstng.mt     0.005    0.004    1.453    0.146
#   Weight.body.age ~~
#     Weight.muscle     0.001    0.001    0.403    0.687
#     Wght.rstng.mt    -0.021    0.030   -0.700    0.484
#   Weight.muscle ~~
#     Wght.rstng.mt     0.007    0.007    1.003    0.316
#
# Variances:
#     Wght.vscrl.ft     0.056    0.006
#     Weight.Omron      2.181    0.245
#     Weight.BMI        0.001    0.000
#     Weight.body.g     0.083    0.009
#     Weight.muscle     0.004    0.000
#     Wght.rstng.mt     1.721    0.193

Comparing the coefficients by eye, they tend to be quite close (usually within 0.1) and the p-values are all statistically-significant.

The network itself looks right, although some of the edges are surprises: I didn’t know visceral fat was predictable from body fat (I thought they were measuring separate things), and the relative independence of muscle suggests that in any exercise plan I might be better off focusing on the body fat percentage rather than the muscle percentage since the former may be effectively determining the latter.

So what did I learn here?

• learning network structure and direction of arrows is hard; even with only 7 variables and n=159 (accurate clean data), the hill-climbing algorithm will learn at least 7 wrong arcs.

• and the derived graphs depend disturbingly heavily on choice of algorithm; I used the hc hill-climbing algorithm (since I’m lazy and didn’t want to specify arrow directions), but when I try out the alternatives like iamb on the same data & blacklist, the found graph looks rather different
• Gaussians are, as always, sensitive to outliers: I was surprised the first graph didn’t show BMI connected to anything, so I took a closer look and found I had miscoded a BMI of 28 as 280!
• bnlearn, while not as hard to use as I expected, could still use usability improvements: I should not need to coerce integer data into exactly equivalent numeric types just because bnlearn doesn’t recognize integers; and blacklisting/whitelisting needs to be more powerful - iteratively generating graphs and manually inspecting and manually blacklisting is tedious and does not scale

• hence, it may make more sense to find a graph using bnlearn and then convert it into simultaneous-equations and manipulate it using more mature SEM libraries

Zeo sleep data

Here I look at my Zeo sleep data; more variables, more complex relations, and more unknown ones, but on the positive side, ~12x more data to work with.

zeo <- read.csv("~/wiki/docs/zeo/gwern-zeodata.csv")
zeo$Sleep.Date <- as.Date(zeo$Sleep.Date, format="%m/%d/%Y")

## convert "05/12/2014 06:45" to "06:45"
zeo$Start.of.Night <- sapply(strsplit(as.character(zeo$Start.of.Night), " "), function(x) { x[2] })
## convert "06:45" to 24300
interval <- function(x) { if (!is.na(x)) { if (grepl(" s",x)) as.integer(sub(" s","",x))
else { y <- unlist(strsplit(x, ":")); as.integer(y[[1]])*60 + as.integer(y[[2]]); }
}
else NA
}
zeo$Start.of.Night <- sapply(zeo$Start.of.Night, interval)
## correct for the switch to new unencrypted firmware in March 2013;
## I don't know why the new firmware subtracts 15 hours
zeo[(zeo$Sleep.Date >= as.Date("2013-03-11")),]$Start.of.Night <- (zeo[(zeo$Sleep.Date >= as.Date("2013-03-11")),]$Start.of.Night + 900) %% (24*60)

## after midnight (24*60=1440), Start.of.Night wraps around to 0, which obscures any trends,
## so we'll map anything before 7AM to time+1440
zeo[zeo$Start.of.Night<420 & !is.na(zeo$Start.of.Night),]$Start.of.Night <- (zeo[zeo$Start.of.Night<420 & !is.na(zeo$Start.of.Night),]$Start.of.Night + (24*60))

zeoSmall <- subset(zeo, select=c(ZQ,Total.Z,Time.to.Z,Time.in.Wake,Time.in.REM,Time.in.Light,Time.in.Deep,Awakenings,Start.of.Night,Morning.Feel))
zeoClean <- na.omit(zeoSmall)
# bnlearn doesn't like the 'integer' class that most of the data-frame is in
zeoClean <- as.data.frame(sapply(zeoClean, as.numeric))

Prior knowledge:

• Start.of.Night is temporally first, and cannot be caused
• Time.to.Z is temporally second, and can be influenced by Start.of.Night (likely a connection between how late I go to bed and how fast I fall asleep) & Time.in.Wake (since if it takes 10 minutes to fall asleep, I must spend ≥10 minutes in wake) but not others
• Morning.Feel is temporally last, and cannot cause anything
• ZQ is a synthetic variable invented by Zeo according to an opaque formula, which cannot cause anything but is determined by others
• Total.Z should be the sum of Time.in.Light, Time.in.REM, and Time.in.Deep
• Awakenings should have an arrow with Time.in.Wake but it’s not clear which way it should run
library(bnlearn)
## after a bunch of iteration, blacklisting arrows which violate the prior knowledge
bl <- data.frame(from=c("Morning.Feel", "ZQ", "ZQ", "ZQ", "ZQ", "ZQ", "ZQ", "Time.in.REM", "Time.in.Light", "Time.in.Deep", "Morning.Feel", "Awakenings", "Time.in.Light", "Morning.Feel", "Morning.Feel","Total.Z", "Time.in.Wake", "Time.to.Z", "Total.Z", "Total.Z", "Total.Z"),
to=c("Start.of.Night", "Total.Z", "Time.in.Wake", "Time.in.REM", "Time.in.Deep", "Morning.Feel","Start.of.Night", "Start.of.Night","Start.of.Night","Start.of.Night", "Time.to.Z", "Time.to.Z", "Time.to.Z", "Total.Z", "Time.in.Wake","Time.to.Z","Time.to.Z", "Start.of.Night", "Time.in.Deep", "Time.in.REM", "Time.in.Light"))

zeo.hc <- hc(zeoClean, blacklist=bl)
zeo.iamb         <- iamb(zeoClean, blacklist=bl)
## problem: undirected arc: Time.in.Deep/Time.in.REM; since hc inferred [Time.in.Deep|Time.in.REM], I'll copy that for iamb:
zeo.iamb <- set.arc(zeo.iamb, from = "Time.in.REM", to = "Time.in.Deep")
zeo.gs <- gs(zeoClean, blacklist=bl)
## same undirected arc:
zeo.gs <- set.arc(zeo.gs, from = "Time.in.REM", to = "Time.in.Deep")

## Bigger is better:
score(zeo.iamb, data=zeoClean)
# [1] -44776.79185
score(zeo.gs, data=zeoClean)
# [1] -44776.79185
score(zeo.hc, data=zeoClean)
# [1] -44557.6952
## hc scores best, so let's look at it:
zeo.hc
#   Bayesian network learned via Score-based methods
#
#   model:
#    [Start.of.Night][Time.to.Z|Start.of.Night][Time.in.Light|Time.to.Z:Start.of.Night]
#    [Time.in.REM|Time.in.Light:Start.of.Night][Time.in.Deep|Time.in.REM:Time.in.Light:Start.of.Night]
#    [Total.Z|Time.in.REM:Time.in.Light:Time.in.Deep][Time.in.Wake|Total.Z:Time.to.Z]
#    [Awakenings|Time.to.Z:Time.in.Wake:Time.in.REM:Time.in.Light:Start.of.Night]
#    [Morning.Feel|Total.Z:Time.to.Z:Time.in.Wake:Time.in.Light:Start.of.Night]
#    [ZQ|Total.Z:Time.in.Wake:Time.in.REM:Time.in.Deep:Awakenings]
#   nodes:                                 10
#   arcs:                                  28
#     undirected arcs:                     0
#     directed arcs:                       28
#   average markov blanket size:           7.40
#   average neighbourhood size:            5.60
#   average branching factor:              2.80
#
#   learning algorithm:                    Hill-Climbing
#   score:                                 BIC (Gauss.)
#   penalization coefficient:              3.614556939
#   tests used in the learning procedure:  281
#   optimized:                             TRUE

plot(zeo.hc)
## https://i.imgur.com/nD3LXND.png

fit <- bn.fit(zeo.hc, zeoClean); fit
#
#   Bayesian network parameters
#
#   Parameters of node ZQ (Gaussian distribution)
#
# Conditional density: ZQ | Total.Z + Time.in.Wake + Time.in.REM + Time.in.Deep + Awakenings
# Coefficients:
#    (Intercept)         Total.Z    Time.in.Wake     Time.in.REM    Time.in.Deep      Awakenings
# -0.12468522173   0.14197043518  -0.07103211437   0.07053271816   0.21121000076  -0.56476256303
# Standard deviation of the residuals: 0.3000223604
#
#   Parameters of node Total.Z (Gaussian distribution)
#
# Conditional density: Total.Z | Time.in.Wake + Start.of.Night
# Coefficients:
#    (Intercept)    Time.in.Wake  Start.of.Night
# 907.6406157850   -0.4479377278   -0.2680771514
# Standard deviation of the residuals: 68.90853885
#
#   Parameters of node Time.to.Z (Gaussian distribution)
#
# Conditional density: Time.to.Z | Start.of.Night
# Coefficients:
#    (Intercept)  Start.of.Night
# -1.02898431407   0.01568450832
# Standard deviation of the residuals: 13.51606719
#
#   Parameters of node Time.in.Wake (Gaussian distribution)
#
# Conditional density: Time.in.Wake | Time.to.Z
# Coefficients:
#   (Intercept)      Time.to.Z
# 14.7433880499   0.3289378711
# Standard deviation of the residuals: 19.0906685
#
#   Parameters of node Time.in.REM (Gaussian distribution)
#
# Conditional density: Time.in.REM | Total.Z + Start.of.Night
# Coefficients:
#      (Intercept)           Total.Z    Start.of.Night
# -120.62442964234     0.37864195651     0.06275760841
# Standard deviation of the residuals: 19.32560757
#
#   Parameters of node Time.in.Light (Gaussian distribution)
#
# Conditional density: Time.in.Light | Total.Z + Time.in.REM + Time.in.Deep
# Coefficients:
#   (Intercept)        Total.Z    Time.in.REM   Time.in.Deep
#  0.6424267863   0.9997862624  -1.0000587988  -1.0001805537
# Standard deviation of the residuals: 0.5002896274
#
#   Parameters of node Time.in.Deep (Gaussian distribution)
#
# Conditional density: Time.in.Deep | Total.Z + Time.in.REM
# Coefficients:
#   (Intercept)        Total.Z    Time.in.REM
# 15.4961459056   0.1283622577  -0.1187382535
# Standard deviation of the residuals: 11.90756843
#
#   Parameters of node Awakenings (Gaussian distribution)
#
# Conditional density: Awakenings | Time.to.Z + Time.in.Wake + Time.in.REM + Time.in.Light + Start.of.Night
# Coefficients:
#     (Intercept)        Time.to.Z     Time.in.Wake      Time.in.REM    Time.in.Light
# -18.41014329148    0.02605164827    0.05736596152    0.02291139969    0.01060661963
#  Start.of.Night
#   0.01129521977
# Standard deviation of the residuals: 2.427868657
#
#   Parameters of node Start.of.Night (Gaussian distribution)
#
# Conditional density: Start.of.Night
# Coefficients:
# (Intercept)
# 1413.382886
# Standard deviation of the residuals: 64.43144125
#
#   Parameters of node Morning.Feel (Gaussian distribution)
#
# Conditional density: Morning.Feel | Total.Z + Time.to.Z + Time.in.Wake + Time.in.Light + Start.of.Night
# Coefficients:
#     (Intercept)          Total.Z        Time.to.Z     Time.in.Wake    Time.in.Light
# -0.924662971061   0.004808652252  -0.010127269154  -0.008636841492  -0.002766602019
#  Start.of.Night
#  0.001672816480
# Standard deviation of the residuals: 0.7104115719

## some issues with big residuals at the extremes in the variables Time.in.Light, Time.in.Wake, and Time.to.Z;
## not sure how to fix those
bn.fit.qqplot(fit)
# https://i.imgur.com/fmP1ca0.png

library(lavaan)
Zeo.model1 <- '
Time.to.Z ~ Start.of.Night
Time.in.Wake ~ Total.Z + Time.to.Z
Awakenings ~ Time.to.Z + Time.in.Wake + Time.in.REM + Time.in.Light + Start.of.Night
Time.in.Light ~ Time.to.Z + Start.of.Night
Time.in.REM ~ Time.in.Light + Start.of.Night
Time.in.Deep ~ Time.in.REM + Time.in.Light + Start.of.Night
Total.Z ~ Time.in.REM + Time.in.Light + Time.in.Deep
ZQ ~ Total.Z + Time.in.Wake + Time.in.REM + Time.in.Deep + Awakenings
Morning.Feel ~ Total.Z + Time.to.Z + Time.in.Wake + Time.in.Light + Start.of.Night
'
Zeo.fit1 <- sem(model = Zeo.model1,  data = zeoClean)
summary(Zeo.fit1)
# lavaan (0.5-16) converged normally after 183 iterations
#
#   Number of observations                          1379
#
#   Estimator                                         ML
#   Minimum Function Test Statistic               22.737
#   Degrees of freedom                                16
#   P-value (Chi-square)                           0.121
#
# Parameter estimates:
#
#   Information                                 Expected
#   Standard Errors                             Standard
#
#                    Estimate  Std.err  Z-value  P(>|z|)
# Regressions:
#   Time.to.Z ~
#     Start.of.Nght     0.016    0.006    2.778    0.005
#   Time.in.Wake ~
#     Total.Z          -0.026    0.007   -3.592    0.000
#     Time.to.Z         0.314    0.038    8.277    0.000
#   Awakenings ~
#     Time.to.Z         0.026    0.005    5.233    0.000
#     Time.in.Wake      0.057    0.003   16.700    0.000
#     Time.in.REM       0.023    0.002   10.107    0.000
#     Time.in.Light     0.011    0.002    6.088    0.000
#     Start.of.Nght     0.011    0.001   10.635    0.000
#   Time.in.Light ~
#     Time.to.Z        -0.348    0.085   -4.121    0.000
#     Start.of.Nght    -0.195    0.018  -10.988    0.000
#   Time.in.REM ~
#     Time.in.Light     0.358    0.018   19.695    0.000
#     Start.of.Nght     0.034    0.013    2.725    0.006
#   Time.in.Deep ~
#     Time.in.REM       0.081    0.012    6.657    0.000
#     Time.in.Light     0.034    0.009    3.713    0.000
#     Start.of.Nght    -0.017    0.006   -3.014    0.003
#   Total.Z ~
#     Time.in.REM       1.000    0.000 2115.859    0.000
#     Time.in.Light     1.000    0.000 2902.045    0.000
#     Time.in.Deep      1.000    0.001  967.322    0.000
#   ZQ ~
#     Total.Z           0.142    0.000  683.980    0.000
#     Time.in.Wake     -0.071    0.000 -155.121    0.000
#     Time.in.REM       0.071    0.000  167.090    0.000
#     Time.in.Deep      0.211    0.001  311.454    0.000
#     Awakenings       -0.565    0.003 -178.407    0.000
#   Morning.Feel ~
#     Total.Z           0.005    0.001    8.488    0.000
#     Time.to.Z        -0.010    0.001   -6.948    0.000
#     Time.in.Wake     -0.009    0.001   -8.592    0.000
#     Time.in.Light    -0.003    0.001   -2.996    0.003
#     Start.of.Nght     0.002    0.000    5.414    0.000

Again no major surprises, but one thing I notice is that ZQ does not seem to connect to Time.in.Light, though Time.in.Light does connect to Morning.Feel; I’ve long suspected that ZQ is a flawed summary and thought it was insufficiently taking into account wakes or something else, so it looks like it’s Time.in.Light specifically which is missing. Start.of.night also is more highly connected than I had expected.

Comparing graphs from the 3 algorithms, they don’t seem to differ as badly as the weight ones did. Is this thanks to the much greater data or the constraints?

Genome sequencing costs

# http://www.genome.gov/sequencingcosts/
# http://www.genome.gov/pages/der/sequencing_costs_apr2014.xls
# converted to CSV & deleted cost per base (less precision); CSV looks like:
# https://dl.dropboxusercontent.com/u/182368464/sequencing_costs_apr2014.csv
## Date, Cost per Genome
## Sep-01,"$95,263,072" ## ... sequencing <- read.csv("sequencing_costs_apr2014.csv") sequencing$Cost.per.Genome <- as.integer(gsub(",", "", sub("\\$", "", as.character(sequencing$Cost.per.Genome))))
# interpret month-years as first of month:
sequencing$Date <- as.Date(paste0("01-", as.character(sequencing$Date)), format="%d-%b-%y")
##         Date Cost.per.Genome
## 1 2001-09-01        95263072
## 2 2002-03-01        70175437
## 3 2002-09-01        61448422
## 4 2003-03-01        53751684
## 5 2003-10-01        40157554
## 6 2004-01-01        28780376

l <- lm(log(Cost.per.Genome) ~ Date, data=sequencing); summary(l)
##
## Coefficients:
##                 Estimate   Std. Error  t value   Pr(>|t|)
## (Intercept) 50.969823683  1.433567932  35.5545 < 2.22e-16
## Date        -0.002689621  0.000101692 -26.4486 < 2.22e-16
##
## Residual standard error: 0.889707 on 45 degrees of freedom
## Multiple R-squared:  0.939559,   Adjusted R-squared:  0.938216
## F-statistic: 699.528 on 1 and 45 DF,  p-value: < 2.22e-16
plot(log(Cost.per.Genome) ~ Date, data=sequencing)
## https://i.imgur.com/3XK8i0h.png
# as expected: linear in log (Moore's law) 2002-2008, sudden drop, return to Moore's law-ish ~December 2011?
# but on the other hand, maybe the post-December 2011 behavior is a continuation of the curve
library(segmented)
# 2 break-points / 3 segments:
piecewise <- segmented(l, seg.Z=~Date, psi=list(Date=c(13970, 16071)))
summary(piecewise)
## Estimated Break-Point(s):
##             Est. St.Err
## psi1.Date 12680 1067.0
## psi2.Date 13200  279.8
##
## t value for the gap-variable(s) V:  0 0 2
##
## Meaningful coefficients of the linear terms:
##                 Estimate   Std. Error  t value   Pr(>|t|)
## (Intercept) 35.841699121  8.975628264  3.99322 0.00026387
## Date        -0.001504431  0.000738358 -2.03754 0.04808491
## U1.Date      0.000679538  0.002057940  0.33020         NA
## U2.Date     -0.002366688  0.001926528 -1.22847         NA
##
## Residual standard error: 0.733558 on 41 degrees of freedom
## Multiple R-Squared: 0.962565,  Adjusted R-squared:   0.958
with(sequencing, plot(Date, log(Cost.per.Genome), pch=16)); plot(piecewise, add=T)
## https://i.imgur.com/HSRqkJO.png
# The first two segments look fine, but the residuals are clearly bad for the third line-segment:
# it undershoots (damaging the second segment's fit), overshoots, then undershoots again. Let's try again with more breakpoints:

lots <- segmented(l, seg.Z=~Date, psi=list(Date=NA), control=seg.control(stop.if.error=FALSE, n.boot=0))
summary(segmented(l, seg.Z=~Date, psi=list(Date=as.Date(c(12310, 12500, 13600, 13750,  14140,  14680,  15010, 15220), origin = "1970-01-01", tz = "EST"))))
# delete every breakpoint below t-value of ~|2.3|, for 3 breakpoints / 4 segments:
piecewise2 <- segmented(l, seg.Z=~Date, psi=list(Date=as.Date(c("2007-08-25","2008-09-18","2010-03-12"))))
with(sequencing, plot(Date, log(Cost.per.Genome), pch=16)); plot(piecewise2, add=T)

# the additional break-point is used up on a better fit in the curve. It looks like an exponential decay/asymptote,
# so let's work on fitting that part of the graph, the post-2007 curve:
sequencingRecent <- sequencing[sequencing$Date>as.Date("2007-10-01"),] lR <- lm(log(Cost.per.Genome) ~ Date, data=sequencingRecent); summary(lR) piecewiseRecent <- segmented(lR, seg.Z=~Date, psi=list(Date=c(14061, 16071))); summary(piecewiseRecent) ## Estimated Break-Point(s): ## Est. St.Err ## psi1.Date 14290 36.31 ## psi2.Date 15290 48.35 ## ## t value for the gap-variable(s) V: 0 0 ## ## Meaningful coefficients of the linear terms: ## Estimate Std. Error t value Pr(>|t|) ## (Intercept) 1.13831e+02 6.65609e+00 17.10182 2.0951e-13 ## Date -7.13247e-03 4.73332e-04 -15.06865 2.2121e-12 ## U1.Date 4.11492e-03 4.94486e-04 8.32161 NA ## U2.Date 2.48613e-03 2.18528e-04 11.37668 NA ## ## Residual standard error: 0.136958 on 20 degrees of freedom ## Multiple R-Squared: 0.995976, Adjusted R-squared: 0.994971 with(sequencingRecent, plot(Date, log(Cost.per.Genome), pch=16)); plot(piecewiseRecent, add=T) lastPiece <- lm(log(Cost.per.Genome) ~ Date, data=sequencingRecent[as.Date(15290, origin = "1970-01-01", tz = "EST")<sequencingRecent$Date,]); summary(lastPiece)
## Coefficients:
##                 Estimate   Std. Error  t value   Pr(>|t|)
## (Intercept) 17.012409648  1.875482507  9.07095 1.7491e-05
## Date        -0.000531621  0.000119056 -4.46528  0.0020963
##
## Residual standard error: 0.0987207 on 8 degrees of freedom
## Multiple R-squared:  0.71366,    Adjusted R-squared:  0.677867
with(sequencingRecent[as.Date(15290, origin = "1970-01-01", tz = "EST")<sequencingRecent$Date,], plot(Date, log(Cost.per.Genome), pch=16)); abline(lastPiece) predictDays <- seq(from=sequencing$Date[1], to=as.Date("2030-12-01"), by="month")
lastPiecePredict <- data.frame(Date = predictDays, Cost.per.Genome=c(sequencing$Cost.per.Genome, rep(NA, 305)), Cost.per.Genome.predicted = exp(predict(lastPiece, newdata = data.frame(Date = predictDays)))) nlmR <- nls(log(Cost.per.Genome) ~ SSasymp(as.integer(Date), Asym, r0, lrc), data=sequencingRecent); summary(nlmR) ## ## Parameters: ## Estimate Std. Error t value Pr(>|t|) ## Asym 7.88908e+00 1.19616e-01 65.95328 <2e-16 ## r0 1.27644e+08 1.07082e+08 1.19203 0.2454 ## lrc -6.72151e+00 5.05221e-02 -133.04110 <2e-16 ## ## Residual standard error: 0.150547 on 23 degrees of freedom with(sequencingRecent, plot(Date, log(Cost.per.Genome))); lines(sequencingRecent$Date, predict(nlmR), col=2)

# side by side:
with(sequencingRecent, plot(Date, log(Cost.per.Genome), pch=16))
lines(sequencingRecent$Date, predict(nlmR), col=3) # as we can see, the 3-piece linear fit and the exponential decay fit identically; # but exponential decay is more parsimonious, IMO, so I prefer that. predictDays <- seq(from=sequencingRecent$Date[1], to=as.Date("2020-12-01"), by="month")
data.frame(Date = predictDays, Cost.per.Genome.predicted = exp(predict(nlmR, newdata = data.frame(Date = predictDays))))

http://www.unz.com/gnxp/the-intel-of-sequencing/#comment-677904 https://biomickwatson.wordpress.com/2015/03/25/the-cost-of-sequencing-is-still-going-down/

Proposal: hand-counting mobile app for more fluid group discussions

Groups use voting for decision-making, but existing vote systems are cumbersome. Hand-raising is faster, but does not scale because hand-counting hands is slow. Advances in machine vision may make it possible for AI to count hands in photos accurately. Combined with a smartphone’s camera, this could yield an app for fast voting in even large groups.

Medium-large (>10 people) groups face a problem in reaching consensus: ballot or pen-and-paper voting is sufficiently slow and clunky that it is too costly to use for anything but the most important discussions. A group is forced to adopt other discussion norms and save a formal vote for only the final decision, and even then the long delay kills a lot of enthusiasm and interest. Voting could be used for many more decisions if it could be faster, and of course all existing group votes would benefit from increased speed. (I am reminded of anime conventions and film festivals where, particularly for short films such as AMVs, one seems to spend more time filling out a ballot & passing them along aisles & the staff painfully counting through each ballot by hand than one actually spends watching the media in question!)

It would be better if voting could be as fluent and easy as simply raising your hand like in a small group such as a classroom - a mechanism which makes it so easy to vote that votes can be held as fast as the alternatives can be spoken aloud and a glance suffices to count (an alert group could vote on 2 or 3 topics in the time it takes to read this sentence). But hand-raising, as great as it is, suffers from the flaw that it does not scale due to the counting problem: a group of 500 people can raise their hands as easily as a group of 50 or 5, but it takes far too long to count ~250 hands: the person counting will quickly tire of the tedium, they will make mistakes counting, and this puts a serious lag on each vote, a lag which increases linearly with the number of voters. (Hands can be easy to approximate if almost everyone votes for or against something, but if consensus is so overwhelming, one doesn’t need to vote in the first place! The hard case of almost-balanced votes is the most important case.)

For comparison, I browsed Amazon looking for highly-rated Energy Star AC models with at least 5000 BTU cooling power and costing $250-$300, picking out the Sunpentown WA-8022S 8000 BTU Window Air Conditioner ($271) with 11.3 EER. (Checking some other entries on Amazon, this is fairly representative on both cost & EER.) Question: what is the electrical savings and hence the payback period of a new AC? The efficiency unit here is the EER or energy efficiency ratio, defined as BTUs (amount of heat being moved by the AC) divided by watts consumed. Here we have ACs with 10.7 EER vs 11.2 EER; I need ~10k BTUs to keep the apartment cool (after fixing a lot of cracks, installing an attic fan and two box fans, putting tin foil over some windows, insulation under a floor etc), so the ACs will use up $EER = \frac{10000}{x \text{watts}}$, and then x = 898 watts and 934 watts respectively. (EER is a lot like miles per gallon/MPG as a measure of efficiency, and shares the same drawbacks: from a cost-perspective, EER/MPG don’t necessarily tell you what you want to know and can be misleading and harder to work with than if efficiency were reported as, say, gallons per mile. As watts per BTU or gallons per mile, it is easy to see that after a certain point, the cost differences have become absolutely small enough that improvements are not worth paying for. Going from 30 gallons of gas to 15 gallons of gas is worth more than going from 3 gallons to 1.5 gallons, even if the relative improvement is the same.) So while operating, the two ACs will use 898 watts vs 934 watts or 0.89kWh vs 0.934kWh to cool; a difference of 36 watts or 0.036kWh. Each kWh costs around$0.09 so the cost-difference is $0.00324 per hour. AC is on May-September (5 months), and on almost all day although it only runs intermittently, so say a third of the day or 8 hours, for a total of 1200 hours of operation. Cost-benefit Thus, then the annual benefit from switching to the new AC with 11.2 EER is $\0.00324 \cdot 8 \cdot 30 \cdot 5 = \3.888$ or$3.9.

The cost is $271 amortized over n years. At$3.9 a year, it will take $\frac{\271}{\3.9}$ annually = 68 years to payback (ignoring breakage and discounting/interest/opportunity-cost). This is not good.

Decision: do not replace.

Discounting

To bring in discounting/interest: For what annual payment (cost-savings) would we be willing to pay the price of a new AC? More specifically, if it costs $271 and has an average payout period of 7 years, then at my usual annual discount rate of 5%, how much must each payout be? $\sum \limits_{t=1}^7 \frac{r}{(1+0.05)^t} \geq \271$ r turns out to be ≥$46.83, which sounds about right. (Discounting penalizes future savings, so r should be greater than $\frac{\271}{7}$ or $39, which it is.)$47 is 12x larger than the estimated savings of $3.9, so the conclusion remains the same. We could also work backward to figure out what EEC would justify an upgrade by treating it as an unknown e and solving for it; let’s say it must payback in 7 years (I doubt average AC lifetime is much longer) at least$271, with the same kWh & usage as before, what must the rival EEC be? as an equation:

$(\frac{\frac{10000}{10.7} - \frac{10000}{e}}{1000} \cdot 0.09 \cdot 8 \cdot 30 \cdot 5) > 47$

and solving,

$e > 20.02$

I am pretty sure there are no ACs with EER>20!

Another way to look at it: if a new good AC costs ~$300 and I expect it to last ~7 years, then that’s an annual cost of$43. The current AC’s total annual cost to run is $1200 \text{hours} \cdot \text{kWhs} \cdot \text{cost per kWh}$ or $(8 \cdot 30 \cdot 5) \cdot 0.934 \cdot 0.09 = \101$. So it’s immediately clear that the energy savings must be huge - half! - before it can hope to justify a new purchase.

Sensitivity analysis

The above analyses were done with point-estimates. It’s only fair to note that there’s a lot of uncertainty lurking in those estimates: $0.09 was just the median of the estimates I found for my state’s electricity rates, the AC might be on 4 or 6 months, the hours per day might be considerably higher (or lower) than my guess of 8 hours, 10.7 & 11.2 EERs are probably best-case estimates and the real efficiencies lower (they’re always lower than nominal), the discount rate may be a percent lower or higher and so minimum savings would be off by as much as$4 in either direction, and so on. It would be good to do a bit of a sensitivity analysis to make sure that this is not being driven by any particular number. (Based on the definition, since it’s using mostly multiplication, the final value should be robust to considerable error in estimating each parameter, but you never know.) Throwing together my intuition for how much uncertainty is in each parameter and modeling most as normals, I can simulate my prior distribution of savings:

set.seed(2015-07-26)
simulate <- function() {
BTUs <- rnorm(1, 10000, 100)
EER_old <- 10.7 - abs(rnorm(1, 0, 0.5)) # half-normals because efficiencies only get worse, not better
EER_new <- 11.2 - abs(rnorm(1, 0, 0.5))
kWh <- rnorm(1, 0.09, 0.01)
dailyUsage <- rnorm(1, 8, 2)
months <- sample (4:6, 1)
minimumSavings <- rnorm(1, 47, 4)

annualNetSavings <- ((((BTUs / EER_old ) - (BTUs / EER_new)) / 1000) * kWh * dailyUsage * 30 * months) - minimumSavings
return(annualNetSavings)
}
sims <- replicate(100000, simulate())
summary(sims)
##        Min.     1st Qu.      Median        Mean     3rd Qu.        Max.
## -70.3666500 -46.2051500 -42.3764100 -42.1133700 -38.3134600  -0.7334517
quantile(sims, p=c(0.025, 0.975))
##        2.5%        97.5%
## -53.59989114 -29.13999204

Under every simulation, a new AC is a net loss. (Since we have no observed data to update our priors with, this is an exercise in probability, not Bayesian inference, and so there is no need to bring in JAGS.)

There are two choices: replace or not. The expected-value of a replacement is $100\% \cdot -\42$ or -$42, and the expected-value of not replacing is $100\% \cdot \0$ or$0; the latter is larger than the former, so we should choose the latter and not replace the old AC.

Hence we can be confident that not getting a new AC really is the right decision.

Some ways of dealing with measurement error

Prompted by a question on LessWrong, some examples of how to analyze noisy measurements in R:

## Create a simulated dataset with known parameters, and then run a ML multilevel model, a ML SEM,
## and a Bayesian multilevel model; with the last, calculate Expected Value of Sample Information (EVSI):

## SIMULATE
set.seed(2015-08-11)
## "There is a variable X, x belongs to [0, 100]."
toplevel <- rnorm(n=1, 50, 25)
## "There are n ways of measuring it, among them A and B are widely used."
## "For any given measurer, the difference between x(A) and x(B) can be up to 20 points."
A <- toplevel + runif(1, min=-10, max=10)
B <- toplevel + runif(1, min=-10, max=10)
c(toplevel, A, B)
# [1] 63.85938385 55.43608379 59.42333264
### the true level of X we wish to recover is '63.85'

## "Between two any measurers, x(A)1 and x(A)2 can differ on average 10 points, likewise with B."
### let's imagine 10 hypothetical points are sample using method A and method B
### assume 'differ on average 10 points' here means something like 'the standard deviation is 10'
A_1 <- rnorm(n=10, mean=A, sd=10)
B_1 <- rnorm(n=10, mean=B, sd=10)

data <- rbind(data.frame(Measurement="A", Y=A_1), data.frame(Measurement="B", Y=B_1)); data
#    Measurement           Y
# 1            A 56.33870025
# 2            A 69.07267213
# 3            A 40.36889573
# 4            A 48.67289213
# 5            A 79.92622603
# 6            A 62.86919410
# 7            A 53.12953462
# 8            A 66.58894990
# 9            A 47.86296948
# 10           A 60.72416003
# 11           B 68.60203507
# 12           B 58.24702007
# 13           B 45.47895879
# 14           B 63.45308935
# 15           B 52.27724328
# 16           B 56.89783535
# 17           B 55.93598486
# 18           B 59.28162022
# 19           B 70.92341777
# 20           B 49.51360373

## MLM

## multi-level model approach:
library(lme4)
mlm <- lmer(Y ~ (1|Measurement), data=data); summary(mlm)
# Random effects:
#  Groups      Name        Variance Std.Dev.
#  Measurement (Intercept)  0.0000  0.000000
#  Residual                95.3333  9.763877
# Number of obs: 20, groups:  Measurement, 2
#
# Fixed effects:
#              Estimate Std. Error  t value
# (Intercept) 58.308250   2.183269 26.70685
confint(mlm)
#                    2.5 %       97.5 %
# .sig01       0.000000000  7.446867736
# .sigma       7.185811525 13.444112087
# (Intercept) 53.402531768 63.213970887

## So we estimate X at 58.3 but it's not inside our confidence interval with such little data. Bad luck?

## SEM

library(lavaan)
X.model <- '        X =~ B + A
A =~ a
B =~ b'
X.fit <- sem(model = X.model, meanstructure = TRUE, data = data2)
summary(X.fit)
# ...                   Estimate  Std.err  Z-value  P(>|z|)
# Latent variables:
#   X =~
#     B                 1.000
#     A              7619.504
#   A =~
#     a                 1.000
#   B =~
#     b                 1.000
#
# Intercepts:
#     a                58.555
#     b                58.061
#     X                 0.000
#     A                 0.000
#     B                 0.000
## Well, that didn't work well - explodes, unfortunately. Probably still not enough data.

## MLM (Bayesian)

library(R2jags)
## rough attempt at writing down an explicit multilevel model which
## respects the mentioned priors about errors being reasonably small:
model <- function() {
grand.mean ~ dunif(0,100)

delta.between.group ~ dunif(0, 10)

sigma.between.group ~ dunif(0, 100)
tau.between.group <- pow(sigma.between.group, -2)

for(j in 1:K){
# let's say the group-level differences are also normally-distributed:
group.delta[j] ~ dnorm(delta.between.group, tau.between.group)
# and each group also has its own standard-deviation, potentially different from the others':
group.within.sigma[j] ~ dunif(0, 20)
group.within.tau[j] <- pow(group.within.sigma[j], -2)

# save the net combo for convenience & interpretability:
group.mean[j] <- grand.mean + group.delta[j]
}

for (i in 1:N) {
# each individual observation is from the grand-mean + group-offset, then normally distributed:
Y[i] ~ dnorm(grand.mean + group.delta[Group[i]], group.within.tau[Group[i]])
}

}
jagsData <- list(N=nrow(data), Y=data$Y, K=length(levels(data$Measurement)),
Group=data$Measurement) params <- c("grand.mean","delta.between.group", "sigma.between.group", "group.delta", "group.mean", "group.within.sigma") k1 <- jags(data=jagsData, parameters.to.save=params, inits=NULL, model.file=model); k1 # ... mu.vect sd.vect 2.5% 25% 50% 75% 97.5% Rhat n.eff # delta.between.group 4.971 2.945 0.221 2.353 4.967 7.594 9.791 1.008 260 # grand.mean 52.477 11.321 23.453 47.914 53.280 58.246 74.080 1.220 20 # group.delta[1] 6.017 11.391 -16.095 0.448 5.316 10.059 34.792 1.152 21 # group.delta[2] 5.662 11.318 -15.836 0.054 5.009 10.107 33.548 1.139 21 # group.mean[1] 58.494 3.765 50.973 56.188 58.459 60.838 66.072 1.001 3000 # group.mean[2] 58.139 2.857 52.687 56.366 58.098 59.851 63.999 1.003 920 # group.within.sigma[1] 12.801 2.766 8.241 10.700 12.446 14.641 18.707 1.002 1100 # group.within.sigma[2] 9.274 2.500 5.688 7.475 8.834 10.539 15.700 1.002 1600 # sigma.between.group 18.031 21.159 0.553 3.793 9.359 23.972 82.604 1.006 1700 # deviance 149.684 2.877 145.953 147.527 149.081 151.213 156.933 1.001 3000 ## VOI posteriorXs <- k1$BUGSoutput$sims.list[["grand.mean"]] MSE <- function(x1, x2) { (x2 - x1)^2 } lossFunction <- function(x, predictions) { mean(sapply(predictions, function(x2) { MSE(x, x2)}))} ## our hypothetical mean-squared loss if we predicted, say, X=60: lossFunction(60, posteriorXs) # [1] 184.7087612 ## of the possible values for X, 1-100, what value of X minimizes our squared error loss? losses <- sapply(c(1:100), function (n) { lossFunction(n, posteriorXs);}) which.min(losses) # [1] 52 ## 52 also equals the mean estimate of X, which is good since it's well known that the mean is what minimizes ## the loss when the loss is squared-error so it suggests that I have not screwed up the definitions losses[52] [1] 128.3478462 ## to calculate EVSI, we repeatedly simulate a few hundred times the existence of a hypothetical 'C' measurement ## and draw n samples from it; ## then we add the C data to our existing A & B data; run our Bayesian multilevel model again on the bigger dataset;, ## calculate what the new loss is, and compare it to the old loss to see how much the new data ## reduced the loss/mean-squared-error. ## Done for each possible n (here, 1-30) and averaged out, this tells us how much 1 additional datapoint is worth, ## 2 additional datapoints, 3 additional datapoints, etc. sampleValues <- NULL for (i in seq(from=1, to=30)) { evsis <- replicate(500, { n <- i C <- toplevel + runif(1, min=-10, max=10) C_1 <- rnorm(n=n, mean=C, sd=10) ## all as before, more or less: newData <- rbind(data, data.frame(Measurement="C", Y=C_1)) jagsData <- list(N=nrow(newData), Y=newData$Y, K=length(levels(newData$Measurement)), Group=newData$Measurement)
params <- c("grand.mean","delta.between.group", "sigma.between.group", "group.delta", "group.mean",
"group.within.sigma")
jEVSI <- jags(data=jagsData, parameters.to.save=params, inits=NULL, model.file=model)

posteriorTimesEVSI <- jEVSI$BUGSoutput$sims.list[["grand.mean"]]
lossesEVSI <- sapply(c(1:100), function (n) { lossFunction(n, posteriorTimesEVSI);})

oldOptimum <- 128.3478462 # losses[52]
newOptimum <- losses[which.min(lossesEVSI)]
EVSI <- newOptimum - oldOptimum
return(EVSI)
}
)

print(i)

print(mean(evsis))
sampleValues[i] <- mean(evsis)
}
sampleValues
#  [1] 13.86568780 11.07101087 14.15645538 13.05296681 11.98902668 13.86866619 13.65059093 14.05991443
#  [9] 14.80018511 16.36944874 15.47624541 15.64710237 15.74060632 14.79901214 13.36776390 15.35179426
# [17] 14.31603459 13.70914727 17.20433606 15.89925289 16.35350861 15.09886204 16.30680175 16.27032067
# [25] 16.30418553 18.84776433 17.86881713 16.65973397 17.04451609 19.17173439

## As expected, the gain in reducing MSE continues increasing as data comes in but with diminishing returns;
## this is probably because in a multilevel model like this, you aren't using the _n_ datapoints to estimate X
## directly so much as you are using them to estimate a much smaller number of latent variables, which are then
## the _n_ used to estimate X. So instead of getting hyperprecise estimates of A/B/C, you need to sample from additional
## groups D/E/F/... Trying to improve your estimate of X by measuring A/B/C many times is like trying to estimate
## IQ precisely by administering a WM test a hundred times.

## If we wanted to compare with alternatives like instead sampling n data points from C and a D, it's easy to modify
## the EVSI loop to do so: generate D <- toplevel + runif(1, min=-10, max=10); D_1 <- rnorm(n=n, mean=D, sd=10)
## and now rbind D_1 in as well. At a guess, after 5-10 samples from the current group, estimates of X will be improved more
## by then sampling from a new group.

## Or the loss function could be made more realistic. It's unlikely one is paid by MSE, and if one adds in how much
## money each sample costs, with a realistic loss function, one could decide exactly how much data is optimal to collect.

## To very precisely estimate X, when our measurements are needed to measure at least 3 latent variables,
## requires much more data than usual.

## In general, we can see the drawbacks and benefits of each approach. A canned MLM
## is very fast to write but doesn't let us include prior information or easily run
## additional analyses like how much additional samples are worth. SEM works poorly
## on small samples but is still easy to write in if we have more complicated
## models of measurement error. A full-blown modeling language like JAGS is quite
## difficult to write in and MCMC is slower than other approaches but handles small
## samples without any errors or problems and offers maximal flexibility in using
## the known prior information and then doing decision-theoretic stuff. Overall for
## this problem, I think JAGS worked out best, but possibly I wasn't using LAVAAN
## right and that's why SEM didn't seem to work well.

Value of Information: clinical prediction instruments for suicide

http://slatestarcodex.com/2015/08/31/magic-markers/#comment-232970

I agree. When criticizing the study for claiming the blood levels added predictive power and it’s not clear they did, this is solely a statistical claim and can be done in a vacuum. But when one then goes on to pan the predictive power of the underlying clinical prediction instruments as useless in all circumstances, based on just the prediction stats:

So when people say We have a blood test to diagnose suicidality with 92% accuracy!, even if it’s true, what they mean is that they have a blood test which, if it comes back positive, there’s still less than 50-50 odds the person involved is suicidal. Okay. Say you’re a psychiatrist. There’s a 48% chance your patient is going to be suicidal in the next year. What are you going to do? Commit her to the hospital? I sure hope not. Ask her some questions, make sure she’s doing okay, watch her kind of closely? You’re a psychiatrist and she’s your depressed patient, you would have been doing that anyway. This blood test is not really actionable. And then remember that this isn’t the blood test we have. We have some clinical prediction instruments that do this…But having a blood test for suicide won’t be very useful, even if it works.

One is implicitly making some strong cost-benefit claims here and stepping from statistics (what are the probabilities?) to decision theory (given these probabilities, how should I act?). They are not identical: no AUC graph will ever tell you if a model’s predictions are useful or not, and there is no universal threshold where 92% specificity/sensitivity is totally useless but 95% would make a difference - these clinical prediction instruments might be useless indeed, but that will depend on costs, base rates, and available actions. (I tried to make this point to Coyne on Twitter earlier but I don’t think he understood what I was getting at & he blew me off.)

Discontinuities come from our actions; our inferences are incremental. There are some contexts where a tiny 1% improvement in AUC might be worth a lot (Wall Street) and there are some contexts where sensitivity or specificity of 99% is still useless because it won’t change your actions at all (I’m currently comparing my riding lawn mower to a robotic lawn mower, and thus far, it doesn’t matter how precise my parameters are, the robotic lawn mowers are, to my disappointment, just too expensive right now). I think p-values have shown us how well arbitrary thresholds work out in practice (and remember where they came from in the first place! decision rules set per problem - Gosset, in optimizing a brewery, did not have the pathologies we have with p<0.05 fetishism.) I also don’t believe your choices are really that restricted: you mean if you were absolutely convinced that your patient was about to commit suicide, there is absolutely nothing you could do besides treat them like any other depressive? That seems unlikely. But whatever, even if commitment is the only alternative, there is still a value to the information provided by a clinical prediction instrument, and we can calculate it, and you should if you want to rule it out as having any value, in the same way that in criticizing a study as weak, it’s better to ignore the p-values and just work out the right posterior and demonstrate directly how little evidence it contains.

Let’s try this as an example, it’s not hard or terribly complex (just tedious). So we have a ward of 100 depressive patients where we are interested in preventing suicide; our prior probability is that 7.5% or ~7 of them will commit suicide. The value of a life has been given a lot of different valuations, but $10 million is a good starting point. Action 1: What are our costs or losses? We could say that we expect a loss of 7.5*$10m or -$75m, and if we stand by and do no treatment or intervention whatsoever, we spend no more money and so the total loss is 0 + 0.075 * 100 * 10,000,000 = -$75,000,000

Action 2:

Let’s say they all stay by default for one week and this costs a net $1000 a day; let’s say further that, since commitment is the mentioned alternative, while committed a suicide attempt will fail. And since we know that suicides are so often spontaneous and major depression comes and goes, a frustrated suicide attempt doesn’t simply mean that they will immediately kill themselves as soon as they get out. This 7% comes from a followup period of a year, so the probability any will attempt suicide in the next week might be 0.075/52 or 0.001442307692. So this gives us our default setup: we have 100 patients staying for 7 days at a net cost of$1000 a day or $700,000 total, and by having them stay, we stop an expected average of 0.14 suicides and thus we prevent an expected loss of 0.14 *$10m = $1,440,000, for a total loss of treatment-cost minus treatment-gain plus remaining-loss:$700,000 - (0.14 * $10m) -$10m * 100 * (0.075-(0.075/52)) = -$74,257,692. Note that this loss is smaller than in the scenario in which we don’t do any commitment at all; since one week of suicide-watch reduced the suicide loss more than it cost, this is not surprising. Specifically, the benefit is: action1 - action2 = gain to switching 75000000 - 74257692 =$742,308

Not fantastic, but it’s in the right order of magnitude (you can’t expect more from a low base-rate event and a treatment with such a low probability of making a difference, after all) so it looks plausible, and it’s still more than zero. We can reject the action of not committing them at all as being inferior to committing them for one week.

Action 3:

What if we were instead choosing between one week and committing them for a full year - thus catching the full 7.5% of suicides during the 1-year followup? Does that work? First, the loss from this course of action:

((100*365.2*1000) - (0 * 10000000) - (10000000 * 100 * (0.075-(0.075/1)))) = -$36,520,000 Since there are no suicides, we avoid the default loss of -$75m, but we still have to spend $36,520,000 to pay for the long-term commitment. However, the benefit to the patients has increased dramatically since we stop so many more suicides: action 2 - action 3 =$35,637,692.31

(We go from a loss of -$74m to a loss of -$36m.) So we see action 3 is even better than action 2 for the patients. Of course, we can’t extrapolate out any further than 1 year, because that’s what our followup number is, and we don’t know how the suicide risk falls after the 1 year point - if it drops to ~0, then further commitment is a terrible idea. So I’m not going to calculate out any further. (Since this is all linear stuff, the predicted benefit will increase smoothly over the year and so there’s no point in calculating out alternatives like 1 month, 3 months, 6 months, 9 months, etc.) What’s that, action 3 is totally infeasible and no one would ever agree to this - the patients would scream their heads off and the health insurance companies would never go for it - even if we could show that long commitments do reduce the suicide rate enough to justify the costs? And, among other things, I’ve oversimplified in assuming the 7% risk is evenly distributed over the year rather than a more plausible distribution like exponentially decreasing from Day 1, so likely commitment stops being a good idea more like month 3 or something? Yeah, you’re probably right, so let’s go back to using action 2’s loss as our current best alternative.

Now, having set out some of the choices available, we can find out how much better information is worth. First, let’s ask what the Expected Value of Perfect Information is: if we were able to take our 100 patients and exactly predict which 7 were depressive and would commit suicide this year in the absence of any intervention, where our choice is between committing them for one week or not at all. Given such information we can eject the 93 who we now know were never a suicide risk, and we hold onto the 7 endangered patients, and we have a new loss of the commitment cost of 7 people for a week vs the prevented loss of the chance they will try to commit suicide that week of this year:

((771000) - (0.14 * 10000000) - (10000000 * 7 * (1-(1/52)))) = -$70,004,846 How much did we gain from our perfect information? About$4m:

74257692 - 70004846 = $4,252,846 (This passes our sanity checks: additional information should never hurt us, so the amount should be >=$0, but we are limited by the intervention to doing very little, so the ceiling should be a low amount compared to the total loss, which this is.)

So as long as the perfect information did not cost us more than $4m or so, we would have net gained from it: we would have been able to focus commitment on the patients at maximal risk. So suppose we had a perfect test which cost$1000 a patient to run, and we wanted to know if the gained information was valuable enough to bother with using this expensive test; the answer in this case is definitely yes: with 100 patients, it’ll cost $100,000 to run the test but it’ll save$4.25m for a net profit of $4.15m. In fact, we would be willing to pay per-patient costs up to$42k, at which point we hit break-even (4252846 / 100).

OK, so that’s perfect information. What about imperfect information? Well, imperfect is a lot like perfect information, just, y’know - less so. Let’s consider this test: with the same prior, a negative on it means the patient now has P=0.007 to commit suicide that year, and a positive means P=0.48, and the sensitivity/specificity at 92%. (Just copying that from OP & ButYouDisagree, since those sound plausible.) So when we run the test on our patients, we find of the 4 possible outcomes:

• 85.1 patients are non-suicidal and the test will not flag them
• 7.4 are non-suicidal but the test will flag them
• 6.9 are suicidal and the test will flag them
• 0.6 are suicidal but the test will not flag them

So if we decide whether to commit or not commit solely based on this test, we will send home 85.1 + 0.6 = 85.7 patients (and indeed 0.6/85.7=0.007), and we will retain the remaining 7.4 + 6.9 = 14.3 patients (and indeed, 6.9/14.3=0.48). So our loss is the wrongly ejected patient of 0.6 suicides plus the cost of committing 14.3 patients (both safe and at-risk) for a week in exchange for the gain of a small chance of stopping the suicide of the 6.9 actually at risk:

(10000000*85.7*0.007) + (14.3*7*1000) + (10000000 * (0.4814.3) (1-(1/52))) = -$73,419,100 How much did we gain from our imperfect information? About$0.8m:

74257692 - 73419100 = $838,592 or$8,385.92 per patient. (This passes our sanity check: greater than $0, but much less than the perfect information. The exact amount may seem lame, but as a fraction of the value of perfect information, it’s not too bad: the test gets us 20% - 838592 / 4252846 - of the way to perfection.) And that’s our answer: the test is not worth$0 - it’s worth $8k. And once you know what the cost of administering the test is, you simply subtract it and now you have the Net Expected Value of Information for this test. (I can’t imagine it costs$8k to administer what this sounds like, so at least in this model, the value is highly likely >$0.) By taking the posterior of the test and integrating all the estimated costs and benefits into a single framework, we can nail down exactly how much value these clinical instruments could deliver if used to guide decision-making. And if you object to some particular parameter or assumption, just build another decision-theory model and estimate the new cost. For example, maybe commitment actually costs, once you take into account all the disruption to lives and other such side-effects, not$1000 but net of $5000 per day, what then? Then the gain halves to$438,192, etc. And if it costs $10000 then the test is worth nothing because you won’t commit anyone ever because it’s just way too expensive, and now you know it’s worth$0; or if commitment is so cheap that it’s more like $100 a day, then the test is also worth$0 because you would just commit everyone (since breakeven is then a suicide probability way below 7%, all the way at ~0.4% which is still below the 0.7% which the test can deliver, so the test result doesn’t matter for deciding whether to commit, so it’s worth $0), or if you adopt a more reasonable value of life like$20m, the value of perfect information shoots up (obviously, since the avoided loss doubles) but the value of imperfect information drops like a stone (since now that one suicidal patient sent home blows away your savings from less committing) and the test becomes worthless; and playing with the formulas, you can figure out the various ranges of assumptions in which the test has positive value and estimate how much it has under particular parameters, and of course if parameters are uncertain, you can cope with that uncertainty by embedding this in a Bayesian model to get posterior distributions of particular parameters incorporating all the uncertainty.

So to sum up: there are no hard thresholds in decision-making and imposing them can cost us better decision-making, so to claim additional information is worthless, more analysis needed, and this analysis must be done with respect to the available actions & their consequences, which even under the somewhat extreme conditions here of very weak interventions & low base-rates, suggests that the value of this information is positive.

Bayesian Model Averaging

## original: "Bayesian model choice via Markov chain Monte Carlo methods" Carlin & Chib 1995 http://stats.ma.ic.ac.uk/~das01/MyWeb/SCBI/Papers/CarlinChib.pdf
## Kobe example & data from: "A tutorial on Bayes factor estimation with the product space method", Lodewyckx et al 2011 http://ejwagenmakers.com/2011/LodewyckxEtAl2011.pdf
## Lodewyckx code can be downloaded after registration & email from http://ppw.kuleuven.be/okp/software/scripts_tut_bfepsm/

## "Table 2: Observed field goals (y) and attempts (n) by Kobe Bryant during the NBA seasons of 1999 to 2006."
Year, y,   n,    y.n
1999, 554, 1183, 0.47
2000, 701, 1510, 0.46
2001, 749, 1597, 0.47
2002, 868, 1924, 0.45
2003, 516, 1178, 0.44
2004, 573, 1324, 0.43
2005, 978, 2173, 0.45
2006, 399,  845, 0.47

library(runjags)
model1<-"model{
# 1) MODEL INDEX
# Model index is 1 or 2.
# Prior probabilities based on argument prior1.
# Posterior probabilities obtained by averaging
# over postr1 and postr2.
M ~ dcat(p[])
p[1] <- prior1
p[2] <- 1-prior1
postr1 <- 2-M
postr2 <- 1-postr1

# 2) MODEL LIKELIHOOD
# For each year, successes are Binomially distributed.
# In M1, the success rate is fixed over years.
# In M2, the success rate is year-specific.
for (i in 1:n.years){
successes[i] ~ dbin(pi[M,i], attempts[i])

pi[1,i] <- pi.fixed
pi[2,i] <- pi.free[i]
}

# 3) MODEL 1 (one single rate)
# The fixed success rate is given a Beta prior and pseudoprior.
# Whether it is a prior or pseudoprior depends on the Model index.
pi.fixed ~ dbeta(alpha.fixed[M],beta.fixed[M])
alpha.fixed[1] <- alpha1.prior
beta.fixed[1] <- beta1.prior
alpha.fixed[2] <- alpha1.pseudo
beta.fixed[2] <- beta1.pseudo

# 4) MODEL 2 (multiple independent rates)
# The year-specific success rate is given a Beta prior and pseudoprior.
# Whether it is a prior or pseudoprior depends on the Model index.
for (i in 1:n.years){
pi.free[i] ~ dbeta(alpha.free[M,i],beta.free[M,i])
alpha.free[2,i] <- alpha2.prior
beta.free[2,i] <- beta2.prior
alpha.free[1,i] <- alpha2.pseudo[i]
beta.free[1,i] <- beta2.pseudo[i]
}
# predictive interval for hypothetical 2007 data in which Kobe makes 1000 shots:
successes.new.1 ~ dbin(pi.fixed, 1000)
successes.new.2 ~ dbin(pi.free[n.years], 1000)

#      success.new.weighted ~ dcat(M
}"
# 'prior1' value from paper
data <- list("prior1"=0.000000007451, "n.years"= length(kobe$Year), "successes"=kobe$y, "attempts"=kobe$n, "alpha1.prior"=1, "beta1.prior"=1, "alpha2.prior"=1, "beta2.prior"=1, "alpha1.pseudo"=1, "beta1.pseudo"=1, "alpha2.pseudo"=rep(1,8), "beta2.pseudo"=rep(1,8) ) # inits <- function() { list(mu=rnorm(1),sd=30,t=as.vector(apply(mailSim,1,mean))) } params <- c("pi.free", "pi.fixed", "postr1", "postr2", "M", "successes.new.1", "successes.new.2") j1 <- run.jags(model=model1, monitor=params, data=data, n.chains=getOption("mc.cores"), method="rjparallel", sample=500000); j1 # JAGS model summary statistics from 4000000 samples (chains = 8; adapt+burnin = 5000): # # Lower95 Median Upper95 Mean SD Mode MCerr MC%ofSD SSeff # pi.free[1] 0.3145 0.46864 0.98709 0.47383 0.11553 -- 0.00041958 0.4 75810 # pi.free[2] 0.10099 0.46447 0.77535 0.47005 0.1154 -- 0.00042169 0.4 74887 # pi.free[3] 0.19415 0.4692 0.86566 0.4741 0.11457 -- 0.00040171 0.4 81342 # pi.free[4] 0.020377 0.45146 0.69697 0.45867 0.11616 -- 0.00042696 0.4 74023 # pi.free[5] 0.024472 0.43846 0.7036 0.44749 0.11757 -- 0.00043352 0.4 73548 # pi.free[6] 0.076795 0.43325 0.74944 0.44318 0.11684 -- 0.00043892 0.4 70863 # pi.free[7] 0.06405 0.45033 0.73614 0.45748 0.11541 -- 0.00041715 0.4 76543 # pi.free[8] 0.30293 0.47267 0.97338 0.47708 0.11506 -- 0.00040938 0.4 79000 # pi.fixed 0.039931 0.45756 0.97903 0.49256 0.26498 -- 0.00099537 0.4 70868 # postr1 0 0 1 0.15601 0.36287 0 0.15113 41.6 6 # postr2 0 1 1 0.84399 0.36287 1 0.15113 41.6 6 # M 1 2 2 1.844 0.36287 2 0.15113 41.6 6 # successes.new.1 0 463 940 492.57 265.28 454 0.99543 0.4 71019 # successes.new.2 300 473 971 477.05 116.03 473 0.4152 0.4 78094 getLogBF <- function(prior0, postr0) { log((postr0/(1-postr0)) / (prior0/(1-prior0))) } getLogBF(0.000000007451, 0.15601) # [1] 17.02669704 ## analytic BF: 18.79; paper's MCMC estimate: 18.80; not sure where I lost 1.8 of the BF. Dealing with all-or-nothing unreliability of data Given two disagreeing polls, one small & imprecise but taken at face-value, and the other large & precise but with a high chance of being totally mistaken, what is the right Bayesian model to update on these two datapoints? I give ABC and MCMC implementations of Bayesian inference on this problem and find that the posterior is bimodal with a mean estimate close to the large unreliable poll’s estimate but with wide credible intervals to cover the mode based on the small reliable poll’s estimate. A question was asked of me: what should one infer if one is given what would be definitive data if one could take it at face value - but one suspects this data might be totally incorrect? An example would be if one wanted to know what fraction of people would answer yes to a particular question, and one had a very small poll (n=10) suggesting 90% say yes, but then one was also given the results from a much larger poll (n=1000) saying 75% responded yes - but this poll was run by untrustworthy people, people that, for whatever reason, you believe might make something up half the time. You should be able to learn something from this unreliable poll, but you can’t learn everything from it because you would be burned half the time. If not for this issue of unreliability, this would be an easy binomial problem: specify a uniform or Jeffreys prior on what percentage of people will say yes, add in the binomial data of 9/10, and look at the posterior. But what do we do with the unreliability joker? Binomial First let’s try the simple case, just updating on a small poll of 9/10. We would expect it to be unimodally peaked around 80-90%, but broad (due to the small sample size) and falling sharply until 100% since being that high is a priori unlikely. MCMC using Bayesian First Aid: ## install.packages("devtools") ## devtools::install_github("rasmusab/bayesian_first_aid") library(BayesianFirstAid) b <- bayes.binom.test(oldData$Yes, oldData$N); b # ...number of successes = 9, number of trials = 10 # Estimated relative frequency of success: # 0.85 # 95% credible interval: # 0.63 0.99 # The relative frequency of success is more than 0.5 by a probability of 0.994 # and less than 0.5 by a probability of 0.006 Which itself is a wrapper around calling out to JAGS doing something like this: library(runjags) model_string <- "model { x ~ dbinom(theta, n) theta ~ dbeta(1, 1) }" model <- autorun.jags(model_string, monitor="theta", data=list(x=oldData$Yes, n=oldData$N)); model # JAGS model summary statistics from 20000 samples (chains = 2; adapt+burnin = 5000): # # Lower95 Median Upper95 Mean SD Mode MCerr MC%ofSD SSeff AC.10 psrf # theta 0.63669 0.85254 0.9944 0.83357 0.10329 -- 0.0007304 0.7 20000 0.011014 1.0004 Here is a simulation-based version of Bayesian inference using ABC: oldData <- data.frame(Yes=9, N=10) simulatePoll <- function(n, pr) { rbinom(1, size=n, p=pr); } poll_abc <- replicate(100000, { # draw from our uniform prior p <- runif(1,min=0,max=1) # simulate a hypothetical poll dataset the same size as our original newData <- data.frame(Yes=simulatePoll(oldData$N, p), N=oldData$N) # were they equal? if so, save sample as part of posterior if (all(oldData == newData)) { return(p) } } ) resultsABC <- unlist(Filter(function(x) {!is.null(x)}, poll_abc)) summary(resultsABC) # Min. 1st Qu. Median Mean 3rd Qu. Max. # 0.3260816 0.7750520 0.8508855 0.8336383 0.9117471 0.9991691 hist(resultsABC) # https://i.imgur.com/fn3XYQW.png They look identical, as they should. Binomial with binary unreliability To implement our more complicated version: the original poll remains the same but we add in the complication of a very large poll which 50% of the time is a true measure of the poll response and 50% of the time is drawn uniformly at random. (So if the true poll response is 90%, then half the time the large poll will yield accurate data like 905/1000 or 890/1000, and the rest it will yield 10/1000 or 400/1000 or 700/1000.) This is different from the more common kinds of measurement-error models where it’s generally assumed that the noisy measurements still have some informativeness to them; here there is none. Specifically, this faux poll has yielded the data not 9/10, but 750/1000. ABC Using ABC again: we generate the reliable small poll as before, and we add in an faux poll where we flip a coin to decide if we are going to return a yes count based on the population parameters or just a random number, then we combine the two datasets and check that it’s identical to the actual data, saving the population probability if it is. oldData2 <- data.frame(Yes=c(9,750), N=c(10,1000)); oldData2 # Yes N # 1 9 10 # 2 750 1000 simulateHonestPoll <- function(n, pr) { rbinom(1, size=n, p=pr); } simulateFauxPoll <- function(n, pr, switchp) { if(sample(c(TRUE, FALSE), 1, prob=c(switchp, 1-switchp))) { rbinom(1, size=n, p=pr); } else { round(runif(1, min=0, max=n)); }} poll_abc <- replicate(1000000, { priorp <- runif(1,min=0,max=1) switch <- 0.5 n1 <- 10 n2 <- 1000 data1 <- data.frame(Yes=simulateHonestPoll(n1, priorp), N=n1) data2 <- data.frame(Yes=simulateFauxPoll(n2, priorp, switch), N=n2) newData <- rbind(data1, data2) if (all(oldData2 == newData)) { return(priorp) } } ) resultsABC <- unlist(Filter(function(x) {!is.null(x)}, poll_abc)) summary(resultsABC) # Min. 1st Qu. Median Mean 3rd Qu. Max. # 0.5256471 0.7427098 0.7584650 0.7860109 0.8133581 0.9765648 hist(resultsABC) # https://i.imgur.com/atMz0jg.png The results are interesting and in this case the summary statistics are misleading: the median is indeed around 75% (as we would expect! since that’s the result of the highly precise poll which has a 50% chance of being the truth) but we see the mean is being pulled away towards the original 90% estimate, and plotting the histogram, bimodality emerges. The posterior reports that there’s still a lot of credibility to the 90% point estimate, but between the original diffuseness of that posterior (leaving a lot of probability to lower responses including, say, 75%) and the high certainty that if accurate the responses will definitely be close to 75%, it winds up peaked at a little higher than 75% (since even if the larger poll is honest, the earlier poll did still find 9/10). So it’s not so much that we think the best estimate of true population rate really is 79% (indeed, the mode is more like 75%, but it could easily be far away from 75% and in the 90%s) as we would need to think more about what we want to do with this posterior before we decide how to summarize it. Mixture ABC is slow and would not scale to more hypothetical polls unless we abandoned exact ABC inference and began using approximate ABC (entirely possible in this case; instead of strict equality between the original and simulated data, we’d instead accept a sample of p if the simulated dataset’s fractions were within, say, 1% of the originals); and the simulation would need to be rewritten anyway. MCMC can handle this if we think of our problem as a mixture model: our problem is that we have poll data drawn from two clusters/distributions - one cluster is the true population distribution of opinion, and the other cluster just spits out noise. We have one observation which we know is from first reliable distribution (the 9/10 poll result), and one observation which we’re not sure which of the two it came from (750/1000), but we do know that the indexing probability for mixing the two distributions is 50%. In JAGS, we write down a model in which dcat flips between 1 and 2 if the cluster is not known, specifying which distribution a sample came from and its theta probability, and then we infer the thetas for both distributions. Of course, we only care about the first distribution’s theta since the second one is noise. library(runjags) model1 <- "model { for (i in 1:N) { y[i] ~ dbinom(theta[i], n[i]) theta[i] <- thetaOfClust[ clust[i] ] clust[i] ~ dcat(pi[]) } pi[1] <- switch[1] pi[2] <- switch[2] thetaOfClust[1] ~ dbeta(1,1) thetaOfClust[2] ~ dunif(0,1) }" j1 <- autorun.jags(model1, monitor=c("theta"), data = list(N=nrow(oldData2), y=oldData2$Yes, n=oldData2$N, switch=c(0.5, 0.5), clust=c(1,NA))); j1 # ... Lower95 Median Upper95 Mean SD Mode MCerr MC%ofSD SSeff AC.10 psrf # theta[1] 0.70582 0.75651 0.97263 0.77926 0.07178 -- 0.001442 2 2478 0.12978 1.0011 # theta[2] 0.72446 0.75078 0.77814 0.75054 0.013646 -- 0.00009649 0.7 20000 0.009458 1 plot(j1) # https://i.imgur.com/EaqR0dD.png Sure enough, we get a good match with the ABC estimate: a mean estimate for the population distribution of 78% with a very wide 95% CI and a clearly bimodal distribution with a huge spike at 75%. Since the MCMC mixture model looks completely different from the imperative simulation-based model, the consistency in estimates & distributions gives me some confidence in the results being right. So we can see how we should update our beliefs - by a perhaps surprising amount towards the unreliable datapoint. The original data was too weak to strongly resist the allure of that highly precise poll. Weakening heuristic? We might try to think of it this way: half the time, the large poll means nothing whatsoever, it contains 0% or no information about the population at all; While the other half of the time, it is exactly what it seems to be and 100% informative; so doesn’t that mean that on average we should treat it as containing half the information we thought it did? And the information is directly based on the sample size: a sample 5x as big contains 5x as much information. So perhaps in this case of all-or-nothing accuracy, we could solve it easily by simply weakening the weight put the unreliable information and shrinking the claimed sample size - instead of treating it as 750 of 1000, treat it as 375/500; and if it had been 75,000 of 100,000, convert it to 37,500 of 50,000. This is a simple and intuitive shortcut, but if we think about what the binomial will return as the unreliable poll increases in size or if we look at the results… switch <- 0.5 oldData3 <- data.frame(Yes=c(9,(750*switch)), N=c(10,(1000*switch))) b2 <- bayes.binom.test(sum(oldData3$Yes), sum(oldData3$N)); b2 # # Bayesian First Aid binomial test # # data: sum(oldData3$Yes) and sum(oldData3$N) # number of successes = 384, number of trials = 510 # Estimated relative frequency of success: # 0.75 # 95% credible interval: # 0.71 0.79 # The relative frequency of success is more than 0.5 by a probability of >0.999 # and less than 0.5 by a probability of <0.001 Unfortunately, this doesn’t work because it doesn’t preserve the bimodal aspect of the posterior, and we get a unimodal distribution ever concentrating on its mean, wiping out the existence of the 0.90 peak. If our untrustworthy poll had instead, say, reported 750,000 out of 1 million, that should only make the peak at 0.75 look like a needle - it should be unable to affect the mass around 0.9, because it doesn’t matter if the data is 100 or 1 million or 1 billion, it still only has a 50% chance of being true. It’s a little hard to see this since the mean frequency of 0.75 is fairly close to the mean of 0.78 from the ABC and we might write this off as approximation error in either the ABC estimate or BFA’s MCMC, but if we look at the 95% CI and note that 0.9 is not inside it or if we plot the posterior (plot(b2)), then the absence of bimodality jumps out. So this trick doesn’t work. Dysgenics power analysis Current dysgenic estimates predict that genotypic IQ in the West are falling at a substantial rate, amounting to around half a standard deviation or more over the past century, by 1. reducing the frequency at which intelligence-increasing genetic variants occur (through natural selection against such variants) and 2. by increasing the number of new and potentially harmful genetic mutations (increasing mutation load). Estimates are produced indirectly by surveying reproductive rates or by trying to show decreases in phenotypic traits associated with intelligence; it would obviously be preferable to examine dysgenic effects directly, by observing decreases in frequencies or increases in mutation load in a large sample of Western genetic information such as SNP arrays or whole-genomes (respectively). Such direct testing of dysgenics hypotheses are becoming increasingly feasible due to the exponential decrease in SNP & whole-genome sequencing costs creating large datasets (some publicly available) and the recent identification of some intelligence genes. It remains unclear how large these datasets must be to overcome sampling error and yield informative estimates of changes in frequencies or mutation load, however; datasets like PGP or SSGAC may still be too small to investigate dysgenics. I considered the effect size estimates and under some simple models derive power calculations & power simulations of how large a dataset would be required to have an 80% chance of detecting a dysgenic effect: to detect the decrease in intelligence SNPs using SNP data, n≥30,000; to detect the increase in mutation load in whole genomes, n≥160 I then compare to available datasets: the effect on SNPs can be detected by a large number of existing proprietary databases, but there are no public databases which will be large enough in the foreseeable future; the effect on mutation load, on the other hand, can be detected using solely the currently publicly available dataset from PGP. So I conclude that while only the proprietary databases can directly test dysgenic theories of selection for the foreseeable future, there is an opportunity to analyze PGP genomes to directly test the dysgenic theory of mutation load. The dysgenics hypothesis argues that due to observed reproductive patterns where the highly educated or intelligent tend to have fewer offspring, genotypic IQ (the upper bound on phenotypic IQs set by genes and the sort of thing measured by a polygenic score). If dysgenics is true, then it is an extremely important phenomenon, as important as many things that get far more attention like lead remediation; but to paraphrase Richard Hamming1, just because a problem is important does not mean it is worth working on or researching or discussing if there is no chance of making progress - if the data is hopelessly compromised by many systematic biases which would cause false positives or if the data is too scanty to overcome random error or analyses so flexible that they could deliver any answer the partisan wishes. Phenotypic data will, in all probability, never allow for a clear & decisive answer to the question of whether dysgenics exists or matters, as long-term comparisons are roughly as credible as noting that global piracy rates have declined while global warming increases, or paracetamol consumption rates have increased in tandem with Alzheimer’s rates; only direct examination of genetics will deliver the decisive answer. It would be nice to have an idea of how much genetic data we would need to overcome random error (and hence, whether it’s possible to make progress in the near future), which we can answer by doing some statistical power analyses. Changes over time in genetics could be due to changes within a particular race or population (for example, in all white Englishmen), or could be due to population movements like one group replacing or migrating or merging into another (population genetics has revealed innumerable complex examples historically). The latter is possible thanks to the increasing availability of ancient DNA, often made public for researchers; so one could observe very long-term trends with cumulatively large effects (implying that small samples may suffice), but this approach has serious issues in interpretation and questions about how comparable intelligence variants may be across groups or throughout human evolution. With the former, there is less concern about interpretation due to greater temporal and ethnic homogeneity - if a GWAS on white northern Europeans in 2013 turns up intelligence variants and produces a useful polygenic score, it will almost certainly work on samples of white northern Europeans in 1900 too - but because the time-scale is so short the effect will be subtler and harder to detect. Nevertheless, a result within a modern population would be much more credible, so we’ll focus on that. How subtle and hard to detect an effect are we talking about here? Woodley 2012 summarizes a number of estimates: Early in the 20th century, negative correlations were observed between intelligence and fertility, which were taken to indicate a dysgenic fertility trend (e.g. Cattell, 1936; Lentz, 1927; Maller, 1933; Sutherland, 1929). Early predictions of the rate of dysgenesis were as high as between 1 and 1.5 IQ points per decade (Cattell, 1937, 1936)…In their study of the relationship between intelligence and both completed and partially completed fertility, van Court and Bean (1985) reported that the relationships were predominantly negative in cohorts born between the years 1912 and 1982…Vining (1982) was the first to have attempted an estimation of the rate of genotypic IQ decline due to dysgenesis with reference to a large national probability cohort of US women aged between 24 and 34 years in 1978. He identified significant negative correlations between fertility and IQ ranging from −.104 to −.221 across categories of sex, age and race, with an estimated genotypic IQ decline of one point a generation. In a 10year follow-up study using the same cohort, Vining (1995) re-examined the relationship between IQ and fertility, now that fertility was complete, finding evidence for a genotypic IQ decline of .5 points per generation. Retherford and Sewell (1988) examined the association between fertility and IQ amongst a sample of 9000 Wisconsin high-school graduates (graduated 1957). They found a selection differential that would have reduced the phenotypic IQ by .81 points per generation under the assumption of equal IQs for parents and children. With an estimate of .4 for the additive heritability of IQ, they calculated a more modest genotypic decline of approximately .33 points. The study of Ree and Earles (1991), which employed the NLSY suggests that once the differential fertility of immigrant groups is taken into consideration, the phenotypic IQ loss amongst the American population may be greater than .8 of a point per generation. Similarly, in summarizing various studies, Herrnstein & Murray (1994) suggest that it would be nearly impossible to make the total [phenotypic IQ decline] come out to less than one point per generation. It might be twice that. (p. 364). Loehlin (1997) found a negative relationship between the fertility of American women aged 35-44 in 1992 and their educational level. By assigning IQ scores to each of six educational levels, Loehlin estimated a dysgenesis rate of .8 points in one generation. Significant contributions to the study of dysgenesis have been made by Lynn, 1996 (see also: 2011) whose book Dysgenics: Genetic deterioration in modern populations provided the first estimates of the magnitude of dysgenesis in Britain over a 90 year period, putting the phenotypic loss at .069 points per year (about 1.7 points a generation assuming a generational length of 25 years). In the same study, Lynn estimated that the genotypic IQ loss was 1.64 points per generation between 1920 and 1940, which reduced to .66 points between 1950 and the present. Subsequent work by Lynn has investigated dysgenesis in other populations. For example Lynn (1999) found evidence for dysgenic fertility amongst those surveyed in the 1994 National Opinion Research Center survey, which encompassed a representative sample of American adults, in the form of negative correlations between the intelligence of adults aged 40+ and the number of children and siblings. Lynn estimates the rate of dysgenesis amongst this cohort at .48 points per generation. In a more recent study, Lynn and van Court (2004) estimated that amongst the most recent US cohort for which fertility can be considered complete (i.e. those born in the years 1940-1949), IQ has declined by .9 points per generation. At the country level, Lynn and Harvey (2008) have found evidence of a global dysgenesis of around .86 points between 1950 and 2000, which is projected to increase to 1.28 points in the period from 2000 to 2050. This projection includes the assumption that 35% of the variance in cross-country IQ differences is due to the influence of genetic factors. A subsequent study by Meisenberg (2009), found that the fertility differential between developed and developing nations has the potential to reduce the phenotypic world population IQ mean by 1.34 points per decade (amounting to a genotypic decline of .47 points per decade assuming Lynn & Harvey’s 35% estimate). This assumes present rates of fertility and pre-reproductive mortality within countries. Meisenberg (2010) and Meisenberg and Kaul (2010) have examined the factors through which intelligence influences reproductive outcomes. They found that amongst the NLSY79 cohort in the United States, the negative correlation between intelligence and fertility is primarily associated with g and is mediated in part by education and income, and to a lesser extent by more liberal gender attitudes. From this Meisenberg has suggested that in the absence of migration and with a constant environment, selection has the potential to reduce the average genotypic IQ of the US population by between .4, .8 and 1.2 points per generation. All of these estimates are genetic selection estimates: indirect estimates inferred from IQ being a heritable trait and then treating it as a natural selection/breeding process, where a trait is selected against based on phenotype and how fast the trait decreases in each succeeding generation depends on how genetic the trait is and how harsh the selection is. So variation in these estimates (quoted estimates per generation range from .3 to 3+) is due to sampling error, differences in populations or time periods, expressing the effect by year or generation, the estimate used for heritability, reliability of IQ estimates, and whether additional genetic effects are taken into account - for example, Woodley et al 2015 finds -.262 points per decade from selection, but in another paper argues that paternal mutation load must be affecting intelligence by ~-0.84 in the general population, giving a total of -1 per decade. Dysgenics effects should be observable by looking at genomes & SNP data with known ages/birth-years and looking for increases in total mutations or decreases in intelligence-causing SNPs, respectively. Selection on SNPs Without formally meta-analyzing all dysgenics studies, a good starting point on the selection effect seems like a genetic selection of 1 point per decade or 0.1 points per year or 0.007 standard deviations per year (or 0.7 standard deviations per century). The most common available genetic data is SNP data, which sequence only the variants most common in the general population; SNP data can look at the effects of genetic selection but will not look at new mutations (since a new mutation would not be common enough to be worth putting onto a SNP chip). Given a large sample of SNP data, a birth year (or age), and a set of binary SNP variables which cause intelligence (coded as 1 for the good variant, 0 for the others), we could formulate this as a multivariate regression: glm(cbind(SNP1, SNP2, ... SNP_N) ~ Year, family=binomial) and see if the year variable has a negative sign (increasing passage of time predicts lower levels of the good genes); if it does, this is evidence for dysgenics. Better yet, given information about the effect size of the SNPs, we could for each person’s SNP sum the net effects and then regress on a single variable, giving more precision rather than looking for independent effects on each SNP: lm(Polygenic_score ~ Year). Again a negative sign on the year variable is evidence for dysgenics. Directional predictions are weak, and in this case we have quantitative predictions of how big the effects should be. Most of the public genomes I looked at seem to have the earliest birthdates in the 1950s or so; genomes can come from any age person (parents can give permission, and sequencing has been done prenatally) so the maximum effect is the difference between 1950 and 2015, which is 65*0.007=0.455 standard deviations (but most genomes will come from intermediate birth-dates, which are less informative about the temporal trend - in the optimal experimental design for measuring a linear trend, half the samples would be from 1950 and the other half from 2015). If the genetic total is going down by 0.455SDs, how much do the frequencies of all the good genes go down? One simple model of genotypic IQ would be to treat it as a large number of alleles of equal binary effect: a binomial sum of n=10,000 1/0 variables with P=50% (population frequency) is reasonable. (For example, GIANT has found a large number of variants for height, and the GCTAs indicate that SNPs explain much more of variance than the top Rietveld hits currently account for; this specific model is loosely inspired by Hsu 2014.) In such a model, the average value of the sum is of course n*p=5000, and the SD is sqrt(n*p*(1-p)) or sqrt(10000*0.5*0.5) or 50. Applying our estimate of dysgenic effect, we would expect the sum to fall by 0.455*50=22.75, so we would be comparing two populations, one with a mean of 5000 and a dysgenic mean of 4977.25. If we were given access to all alleles from a sample of 1950 and 2015 genomes and so we could construct the sum, how hard would it be able to tell the difference? In this case, the sum is normally distributed as there are more than enough alleles to create normality, so we can just treat this as a two-sample normally-distributed comparison of means (a t-test), and we already have a directional effect size in mind, -0.445SDs, so: power.t.test(delta=0.455, power=0.8, alternative="one.sided") # Two-sample t test power calculation # # n = 60.4155602 # ... A total n=120 is doable, but it is unlikely that we will know all intelligence genes anytime soon; instead, we know a few. A new mean of 4977 implies that since total number of alleles is the same but the mean has fallen, the frequencies must also fall and the average frequency falls from 0.5 to 4977.25/10000=0.497725. To go to the other extreme, if we know only a single gene and we want to test a fall from a frequency of 0.50 to 0.4977, we need infeasibly more samples: power.prop.test(p1=0.5, p2=0.497725, power=0.8, alternative="one.sided") # Two-sample comparison of proportions power calculation # # n = 597,272.2524 # ... 1.2m datapoints would be difficult to get, and so a single gene test would be unhelpful; further, a single gene could change frequencies solely through genetic drift without the change being due to dysgenic pressures. We know a number of genes, though: Rietveld gives 4 good hits, so we can look at a polygenic score from that. They are all of similar effect size and frequency, so we’ll continue under the same assumptions of 1/0 and P=50%. The non-dysgenic average score is 4*0.5=2, sd=sqrt(4*0.5*0.5)=1. (Naturally, the SD is much larger than before because with so few random variables…) The predicted shift is from frequencies of 0.5 to 0.497, so the dysgenic scores should be 4*0.497=1.988, sd=sqrt(4*0.497*0.503)=0.999. The difference of 0.012 on the reduced polygenic score is d=((2-1.988) / 0.999)=0.012, giving a necessary power of: power.t.test(delta=0.012006003, power=0.8) # Two-sample t test power calculation # # n = 108904.194 # ... So the 4 hits do reduce the necessary sample size, but it’s still not feasible to require 218k SNP datasets (unless you are 23andMe or SSGAC or an entity like that). In the current GWAS literature, there are ~9 hits we could use, but the upcoming SSGAC paper promises: We identified 86 independent SNPs associated with EA (p<5E-8).. So how much would 86 improve over 4? • mean old: 86*0.5=43 • sd old: sqrt(86*0.5*0.5)=4.6368 • mean new: 86*0.497=42.742 • sd new: sqrt(86*0.497*(1-0.497))=4.6367 • so d=(43-42.742)/4.63675=0.0556 power.t.test(delta=((43-42.742)/4.63675), power=0.8) # Two-sample t test power calculation # # n = 5071.166739 # ... So with 75, it drops from 200k to 10.1k. To work backwards: we know with 1 hit, we need a million SNP datasets (infeasible for any but the largest proprietary databases, who have no interest in studying this hypothesis), and with all hits we need more like 200 genomes (entirely doable with just publicly available datasets like PGP), but how many hits do we need to work with an in-between amount of data like the ~2k genomes with ages I guess may be publicly available now or in the near future? power.t.test(n=1000, power=0.8) # Two-sample t test power calculation # # n = 1000 # delta = 0.1253508704 hits=437; mean1=hits*0.5; sd1=sqrt(hits*0.5*0.5); mean2=hits*0.497; sd2=sqrt(hits*0.497*(1-0.497)); d=(mean1-mean2)/mean(c(sd1,sd2)); d # [1] 0.1254283986 With a polygenic score drawing on 437 hits, then a sample of 2k suffices to detect the maximum decrease. This is pessimistic because the 10k alleles are not all the same effect size and GWAS studies inherently will tend to find the largest effects first. So the first 4 (or 86) hits are worth the most. The distribution of effects is probably something like an inverse exponential distribution: many small near-zero effects and a few large ones. Rietveld 2013 released the betas for all SNPs, and the beta estimates can be plotted; each estimate is imprecise and there are artifacts in the beta sizes (SSGAC confirms that they were rounded to 3 decimals), but the distribution looks like a radioactive half-life graph, an inverse exponential distribution. With a mean of 1, we can simulate creating a set of 10k effect sizes which are exponentially distributed and have mean 5000 and SD close to (but larger than) 50 and mimics closely the binomial model: effects <- sort(rexp(10000)/1, decreasing=TRUE) genomeOld <- function() { ifelse(sample(c(FALSE,TRUE), prob=c(0.5, 0.5), 10000, replace = TRUE), 0, effects) } mean(replicate(10000, sum(genomeOld()))) # [1] 5000.270218 sd(replicate(10000, sum(genomeOld()))) # [1] 69.82652816 genomeNew <- function() { ifelse(sample(c(FALSE,TRUE), prob=c(0.497, 1-0.497), 10000, replace = TRUE), 0, effects) } With a dysgenic effect of -0.445SDs, that’s a fall of the sum of random exponentials of ~31, which agrees closely with the difference in polygenic genome scores: mean(replicate(10000, sum(genomeOld() - genomeNew()))) # [1] 29.75354558 For each draw from the old and new populations, we can take the first 4 alleles, which were the ones assigned the largest effects, and build a weak polygenic score and compare means. For example: polyNew <- replicate(1000, sum(genomeNew()[1:4])) polyOld <- replicate(1000, sum(genomeOld()[1:4])) t.test(polyOld, polyNew, alternative="greater") # Welch Two Sample t-test # # data: polyOld and polyNew # t = 0.12808985, df = 1995.8371, p-value = 0.8980908 # alternative hypothesis: true difference in means is not equal to 0 # 95 percent confidence interval: # -0.7044731204 0.8029267301 # sample estimates: # mean of x mean of y # 17.72741040 17.67818359 Or to mimic 86 hits: t.test(replicate(1000, sum(genomeOld()[1:86])), replicate(1000, sum(genomeNew()[1:86]))) # # Welch Two Sample t-test # # t = 1.2268929, df = 1997.6307, p-value = 0.2200074 # alternative hypothesis: true difference in means is not equal to 0 # 95% confidence interval: # -0.8642674547 3.7525210076 # sample estimates: # mean of x mean of y # 244.5471658 243.1030390 Using the exponential simulation, we can do a parallelized power analysis: simulate draws (i=300) & tests for a variety of sample sizes to get an idea of what sample size we need to get decent power with 86 hits. library(ggplot2) library(parallel) # warning, Windows users library(plyr) genomeOld <- function(efft) { ifelse(sample(c(FALSE,TRUE), prob=c(0.5, 0.5), length(efft), replace = TRUE), 0, efft) } genomeNew <- function(efft) { ifelse(sample(c(FALSE,TRUE), prob=c(0.497, 1-0.497), length(efft), replace = TRUE), 0, efft) } simulateStudy <- function(n, hits) { effects <- sort(rexp(10000)/1, decreasing=TRUE)[1:hits] polyOld <- replicate(n, sum(genomeOld(effects))) polyNew <- replicate(n, sum(genomeNew(effects))) t <- t.test(polyOld, polyNew, alternative="greater") return(data.frame(N=n, P=t$p.value, PO.mean=mean(polyOld), PO.sd=sd(polyOld), PN.mean=mean(polyNew), PN.sd=sd(polyNew))) }

hits <- 86
parallelStudies <- function(n, itr) { ldply(mclapply(1:itr, function(x) { simulateStudy(n, hits); })); }

sampleSizes <- seq(500, 5000, by=100)
iters <- 300
powerExponential <- ldply(lapply(sampleSizes, function(n) { parallelStudies(n, iters) })); summary(powerExponential)
#        N              P                  PO.mean             PO.sd             PN.mean
#  Min.   : 500   Min.   :0.000000000   Min.   :222.5525   Min.   :23.84966   Min.   :221.2894
#  1st Qu.:1600   1st Qu.:0.002991554   1st Qu.:242.8170   1st Qu.:26.46606   1st Qu.:241.3242
#  Median :2750   Median :0.023639517   Median :247.2059   Median :27.04467   Median :245.7044
#  Mean   :2750   Mean   :0.093184735   Mean   :247.3352   Mean   :27.06300   Mean   :245.8298
#  3rd Qu.:3900   3rd Qu.:0.107997575   3rd Qu.:251.7787   3rd Qu.:27.64103   3rd Qu.:250.2157
#  Max.   :5000   Max.   :0.997322161   Max.   :276.2614   Max.   :30.67000   Max.   :275.7741
#      PN.sd
#  Min.   :23.04527
#  1st Qu.:26.45508
#  Median :27.04299
#  Mean   :27.05750
#  3rd Qu.:27.63241
#  Max.   :30.85065
powerExponential$Power <- powerExponential$P<0.05
powers <- aggregate(Power ~ N, mean, data=powerExponential); powers
# 1   500 0.2133333333
# 2   600 0.2833333333
# 3   700 0.2833333333
# 4   800 0.3133333333
# 5   900 0.3033333333
# 6  1000 0.3400000000
# 7  1100 0.4066666667
# 8  1200 0.3833333333
# 9  1300 0.4133333333
# 10 1400 0.4166666667
# 11 1500 0.4700000000
# 12 1600 0.4600000000
# 13 1700 0.4666666667
# 14 1800 0.4733333333
# 15 1900 0.5233333333
# 16 2000 0.5366666667
# 17 2100 0.6000000000
# 18 2200 0.5900000000
# 19 2300 0.5600000000
# 20 2400 0.6066666667
# 21 2500 0.6066666667
# 22 2600 0.6700000000
# 23 2700 0.6566666667
# 24 2800 0.7133333333
# 25 2900 0.7200000000
# 26 3000 0.7300000000
# 27 3100 0.7300000000
# 28 3200 0.7066666667
# 29 3300 0.7433333333
# 30 3400 0.7133333333
# 31 3500 0.7233333333
# 32 3600 0.7200000000
# 33 3700 0.7766666667
# 34 3800 0.7933333333
# 35 3900 0.7700000000
# 36 4000 0.8100000000
# 37 4100 0.7766666667
# 38 4200 0.8000000000
# 39 4300 0.8333333333
# 40 4400 0.8466666667
# 41 4500 0.8700000000
# 42 4600 0.8633333333
# 43 4700 0.8166666667
# 44 4800 0.8366666667
# 45 4900 0.8666666667
# 46 5000 0.8800000000
qplot(N, Power, data=powers)  + stat_smooth()

So for a well-powered two-group comparison of 1950 & 2015 SNP datasets using 86 SNPs, we would want ~4000 in each group for a total n=8000; we do have nontrivial power even at a total n=1000 (500 in each group means 21% power) but a non-statistically-significant result will be difficult to interpret and if one wanted to do that, reporting a Bayes factor from a Bayesian hypothesis test would make much more sense to express clearly whether the (non-definitive) data is evidence for or against dysgenics.

This is still too optimistic since we assumed the optimal scenario of only very old and very new genomes, while available genomes are more likely to be distributed fairly uniformly between 1950 and 2015. Per Optimal design in psychological research, McClelland 1997, we expect a penalty of ~2x in sample size efficiency in going from the optimal two-group extreme endpoints design to samples being uniformly distributed (due to much of our sample size being wasted on estimating small effects) and so we would expect our sample size requirement to at least double to around n=16000, but we can do a power simulation here as well. To get the effect size for each year, we simply split the frequency decrease over each year and generate hypothetical genomes with less of a frequency decrease uniformly distributed 1950-2015, and do a linear regression to get a p-value for the year predictor:

hits <- 86
sampleSizes <- seq(8000, 30000, by=1000)
iters <- 100
genome <- function(effects) {
t <- sample(c(1:(2015-1950)), 1)
decreasedFrequency <- 0.5 - (((0.5-0.497)/(2015-1950)) * t)
geneFlips <- sample(c(FALSE,TRUE), prob=c(decreasedFrequency, 1-decreasedFrequency), replace = TRUE, length(effects))
geneValues <- ifelse(geneFlips, effects, 0)
return(data.frame(Year=1950+t,
PolygenicScore=sum(geneValues)))
}
simulateStudy <- function(n, hits) {
effects <- sort(rexp(10000)/1, decreasing=TRUE)[1:hits]
d <- ldply(replicate(n, genome(effects), simplify=FALSE))
l <- lm(PolygenicScore ~ Year, data=d)
p <- anova(l)$Pr(>F)[1] return(data.frame(N=n, P=p, PO.mean=predict(l, newdata=data.frame(Year=1950)), PN.mean=predict(l, newdata=data.frame(Year=2015)))) } parallelStudies <- function(n, itr) { ldply(mclapply(1:itr, function(x) { simulateStudy(n, hits); })); } powerExponentialDistributed <- ldply(lapply(sampleSizes, function(n) { parallelStudies(n, iters) })); summary(powerExponential) powerExponentialDistributed$Power <- powerExponentialDistributed$P<0.05 powers <- aggregate(Power ~ N, mean, data=powerExponentialDistributed); powers # N Power # 1 8000 0.27 # 2 9000 0.32 # 3 10000 0.35 # 4 11000 0.33 # 5 12000 0.41 # 6 13000 0.34 # 7 14000 0.41 # 8 15000 0.48 # 9 16000 0.55 # 10 17000 0.62 # 11 18000 0.55 # 12 19000 0.60 # 13 20000 0.69 # 14 21000 0.61 # 15 22000 0.65 # 16 23000 0.63 # 17 24000 0.71 # 18 25000 0.67 # 19 26000 0.71 # 20 27000 0.74 # 21 28000 0.70 # 22 29000 0.79 # 23 30000 0.83 qplot(N, Power, data=powers) + stat_smooth() In this case, the power simulation suggestions the need for triple rather than double the data, and so a total of n=30,000 to be well-powered. Mutation load The paternal mutation load should show up as a increase (70 new mutations per generation, 35 years per generation, so ~2 per year on average) over the past century, while the genetic selection will operate by reducing the frequency of variants which increase intelligence. If there are ~70 new mutations per generation and 2 harmful, and there is no longer any purifying selection so that all 70 will tend to remain present, how much does that compare to existing mutation load averages and, more importantly, standard deviations? A mutation load review leads me to some hard figures from Simons et al 2014 (supplement) using data from Fu et al 2012; particularly relevant is figure 3, the number of single-nucleotide variants per person over the European-American sample, split by estimates of harm from least to most likely: 21345 + 15231 + 5338 + 1682 + 1969 = 45565. The supplementary tables gives a count of all observed SNVs by category, which sum to 300209 + 8355 + 220391 + 7001 + 351265 + 10293 = 897514, so the average frequency must be 45565/897514=0.05, and then the binomial SD will be sqrt(897514*0.05*(1-0.05))=206.47. Considering the two-sample case of 1950 vs 2015, that’s an increase of 130 total SNVs (65*2), which is 0.63SDs, hence: power.t.test(d=(130/206), power=0.8) # Two-sample t test power calculation # # n = 40.40035398 # ... A total of n=80. This particular set up for the two-sample test can be seen as a linear model with the optimum design of allocating half the sample to each extreme (see again McClelland 1997); but more realistically, there is an even distribution across years, in which case the penalty is 2x and n=160. Weaknesses There are some potential problems: 1. Range restriction: in many IQ-related studies, failure to account for selection effects yielding a limited range of IQs may seriously understate the true correlation; this is true in general but particularly common in IQ studies because selection on IQ (eg samples of convenience using only college students) is so universal in human society This may not be such a large issue when dealing with polygenic scores; even severe IQ selection effects will increase polygenic scores only somewhat because the polygenic scores explain so little of IQ variance in the first place. 2. Self-selection by age: if people providing genetic data are not random samples, then there may be pseudo-trends which can mask a real dysgenic trend or create a pseudo-dysgenic trend where there is none. For example, if young people buying genome or SNP data tend to be above-average in intelligence and scientific interest (which anecdotally they certainly do seem to be), while old people tend to get genomes or SNP data due to health problems (and otherwise have average levels of intelligence and thus polygenic score), then in comparing young vs old, one might find not a dysgenic but a pseudo-eugenic trend instead! Conversely, it could be the other way around, if much fewer elderly get genetic data and younger people are more concerned about future health or are going along with a fad, producing a pseudo-dysgenic effect instead (eg in the PGP genome data, there seem to be disproportionately more PhDs who are quite elderly, while younger participants are a more scattershot sample from the general population; probably relating to the circumstances of PGP’s founding & Harvard home). This is probably an issue with databases that rely on voluntary individual contributions, such as PGP, where selection effects have free play. It would be much less of an issue with longitudinal studies where motivations and participation rates will not differ much by age. Since most dysgenic theories accept that recorded IQ scores have remained stable over the 20th century and the decreases in genetic potential either have not manifested yet or have been masked by the Flynn effect & greater familiarity with tests & loss of some g-loading, one might reason that proxies like educational achievement should be increasing throughout one’s sample (since they are known to have increased), and a lack of such a trend indicates selection bias. Genetic data availability Proprietary The known proprietary databases have long been large enough to carry out either analysis, as well as countless other analyses (but have failed to and represent a tragedy of the anticommons): 1. The mutation load analysis requires a whole-genome sample size small enough to have been carried out by innumerable groups post-2009. 2. For SNPs, an incomplete list of examples of publications based on large samples: • 23andMe reached 1 million customers in July 2015, of whom ~80% opt-in to research; the first questions 23andMe asks all customers are age and education, so they likely have at least 700,000 usable SNPs for both discovering educational associations & dysgenic tests. In June 2010, they claimed to have 29k opt-ins out of 50k customers, implying they were well-powered for a dysgenic test in 2010 if they had access to a polygenic score, and that in the absence of the score, they could have found the 85 SSGAC hits (using n=305k) themselves somewhere around mid-2011 or 2012 and then done a dysgenic test. • the SSGAC collaboration has n=305k as of late 2015 • GIANT: height: n=253,288 • cholesterol: n=188,000 • UK Biobank: n=152,729 (sequenced/published on as of June 2015; 500k were enrolled and will be covered eventually) • diabetes: n=114,981 • Psychiatric Genomics Consortium: has run multiple studies of varying sizes, the second-largest (a bipolar study) peaking at a total control group of n=51672, and the largest (schizophrenia) at a control group of n=113,075 • Parkinson’s: n=100,833 (may overlap with 23andMe and some others) • eczema: n=95,464 • Genetics of Personality Consortium: eg Neuroticism: n=63k • Dutch LifeLines Biobank and Cohort: n>13,000 • Health and Retirement Survey: n=12,500 • Swedish TwinGene project: n=10,682 • TwinsUK registry: n=4,905 • Generation Scotland: the Scottish Family Health Study: n=24,000 The existing private groups do not seem to have any interest in testing dysgenics, with the possible exception of future GWAS studies examining fertility, one of which is mentioned by Mills & Tropf 2015: At the time of writing this review, Mills and her research team at the University of Oxford are currently leading a large consortium to engage in the first ever genome-wide association search (GWAS) and meta-analysis of reproductive choice (age at first birth; number of children), conducted in both men and women in over 50 data sets, with the results replicated in additional datasets in a large sample. The hits in such a GWAS might overlap with intelligence hits, and if the multiple hits increase intelligence but decrease fertility or vice versa (as compared to decreasing or increasing both), that would be evidence for dysgenics. Or, assuming the betas are reported, polygenic scores for fertility and intelligence could be estimated in independent samples and checked for an inverse correlation. Public There are a few sources of data, primarily SNP data, which are freely available to all users: 1. 1000 Genomes: unusable due to a deliberate policy decision by 1000 Genomes to delete all phenotype data, including age; similar is 69 Genomes. Both likely would be unusable due to the diversity of the global sample (there is no reason to think that dysgenics pressures are operating in every population at the same strength) 2. OpenSNP: hosting for user-provided SNP & phenotype data with dumps available; hosts ~2k SNP datasets, but only 270 users have birth-years 3. SNPedia likewise hosts SNP data (overlapping with OpenSNP) and genome data, but a very small number 4. Genomes unzipped provides a small amount of data 5. DNA.LAND: claims n=8k based on public participation & input, but seems to then restrict access to a small set of researchers 6. Exome Aggregation Consortium: n=61,486 exomes; phenotype data is unavailable 7. Personal Genome Project (PGP): probably the single largest source of open SNP & genome data. ~1252 participants have registered birthdates according to demographics.tsv, and their statistics page’s graphs indicates <300 whole genomes and <1k SNPs. Phenotype data has been recently released as a SQLite database, making it easier to work with. • Genomes: browsing the user lists for Whole genome datasets, I estimate a total of ~222; looking at the first and last 22 entries, 34 had ages/birth-years, so ~75% of the whole genomes come with the necessary birth-year data, indicating ~166 usable genomes for the purpose of testing dysgenics. With the most recent one uploaded on 2015-10-12, and the earliest recorded being 2011-09-16, that suggests the available genome number increases by ~0.25/day. 166 is uncomfortably close to the requirement for a well-powered test, and there may not be enough data to account for glitches in the data or allow for more complicated statistical testing, but if we wanted to double the available data, we’d only need to wait around 885 days or 2.5 years (or less, depending on whether the collapse in genome sequencing prices continue and prices drop below even the current$1k genomes).
• SNPs: PGP has ~656 23andMe SNP datasets (the number of SNP datasets sourced from other providers is quite small so I didn’t include them), dated 2015-10-21-2011-01-06, so assuming same birth-date percentage, 0.37 per day. Unfortunately, to get 30k SNP datasets through PGP, we would have to wait (linearly extrapolating) 291 years. (Making matters worse, in October 2015, 23andMe doubled its price and reduced the quality of SNP coverage, which will discourage many users and push other users to purchase whole-genome sequencing instead.)

Power analysis for racial admixture studies of continuous variables

I consider power analysis of a genomic racial admixture study for detecting genetic group differences affecting a continuous trait such as IQ in US African-Americans, where ancestry is directly measured by genome sequencing and the comparisons are all within-family to eliminate confounding by population structure or racism/colorism/discrimination. The necessary sample size for well-powered studies is closely related to the average size of differences in ancestry percentage between siblings, as the upper bound on IQ effect per percentage is small, requiring large differences in ancestry to detect easily. A within-family comparison of siblings, due to the relatively small differences in ancestry between siblings estimated from IBD measurements of siblings, might require n>50,000 pairs of siblings to detect possible effects on IQ, an infeasible sample size. An alternative design focuses on increasing the available ancestry differences within a family unit by comparing adoptees with siblings; the larger within-population standard deviation of ancestry creates larger & more easily-detected IQ differences. A random-effects meta-analysis of past admixture & ancestry studies suggests the SD in heterogeneous samples may range from 2% to 20% with a mean of 11% (95% predictive interval), yielding sample sizes of n>20,000, n=1100, and n=500. Hence, an adoption study is probably in the feasible range, with required sample sizes comparable to annual adoption rates among US African-Americans.

Admixture studies examine racial phenotypic differences in traits such as blood pressure by comparing people with ancestry from multiple groups, and correlating differences in ancestry percentage with differences in the phenotype. So, for example, African-Americans have higher blood-pressure than white Americans, and most African-Americans have an average white ancestry of something like 20-25% (see later); if having 26% white ancestry predicts slightly lower blood pressure while 24% predicts higher, that suggests the difference is (as is currently believed) genetic; and this logic can be used to narrow down to specific chromosome regions, and has contributed to study of racial differences in disease.

One application would be to thorny questions like potential group differences in non-medical traits like intelligence. The standard admixture design, requiring a few thousand subjects spanning the full range, might not necessarily work here here because of the claimed environmental effects. A proposed resolution to the question is to do an admixture study comparing African-American siblings. Siblings are highly genetically related on average (50%) but in a randomized fashion due to recombination; so two siblings, including fraternal twins, born to the same parents in the same family in the same neighborhood going to the same schools, will nevertheless have many different variants, and will differ in how related they are - the average is 50% but it could be as low as 45% or high as 55%. So given two siblings, they will differ slightly in their white ancestry, and if indeed white ancestry brings with it more intelligence variants, then the sibling with a higher whiter percentage ought to be slightly more intelligent on average, and this effect will have to be causal, as the inheritance is randomized and all other factors are equal by design. (A result using ancestry percentages measured in the general population, outside families, would be able to make far more powerful comparisons by comparing people with ~0% white ancestry to those with anywhere up to 100%, and require small sample sizes, and such analyses have been done with the expected result, but are ambiguous & totally unconvincing, as the correlation of greater whiteness with intelligence could easily be due to greater SES or greater blackness could be a marker for recent immigration or any of a number of confounds that exist.) This has historically been difficult or impossible since how does one measure the actual ancestry in siblings? But with the rise of cheap genotyping, precise measure of actual (rather than average) ancestry can be done for <$100, so that is no longer an obstacle. Sibling power analysis How many sibling pairs would this require? • you are trying to regress IQ_difference ~ Ancestry_difference • the SD of the IQ difference of siblings is known - it’s ~13 IQ points (nonshared environment + differences in genetics) • of this, a small fraction will be explained by the small difference in ancestry percentage • the power will be determined by the ratio of the sibling SD to the IQ-difference-due-to-ancestry-difference SD, giving an effect size, which combined with the usual alpha=0.05 and beta=0.80, uniquely determines the sample size • IQ-difference-due-to-ancestry-difference SD will be the advantage of better ancestry times how much ancestry differs • if you knew the number of relevant alleles, you could calculate through the binomial the expected SD of sibling ancestor differences. As there are so many alleles, it will be almost exactly normal. So it’s not surprising that siblings overall, for all variants, are 50% IBD with a SD of 4%. If we treated it as simply as possible, Visscher 2006 for an analogous height analysis says they measured 588 markers. So a binomial with 588 draws and p=0.5 implies that 147 markers are expected to be the same: 588 * 0.5*(1-0.5) # [1] 147 and the distribution around 147 is 12, which is ~8%: sqrt((588 * 0.5*(1-0.5))) # [1] 12.12435565 12/147 # [1] 0.08163265306 Visscher does a more complicated analysis taking into account closeness of the markers and gets a SD of 3.9%: equation 7; variance = 1/(16*L) - (1/3*L^2), where L = 35, so L=35; sqrt(1/(16*L) - (1/(3*L^2))) # [1] 0.03890508247 So whatever the mean admixture is, I suppose it’ll have a similar SD of 4-8% of itself. IIRC, African-Americans are ~25% admixed, so with a mean admixture of 25%, we would expect siblings differences to be 25% +- 0.04*0.25=0.01 or 1% difference. If that 75% missing white ancestry accounts for 9 IQ points or 0.6SDs, then each percentage of white ancestry would be 0.6/75 =0.008 SDs. So that SD of 1% more white ancestry yields an SD of 0.008 IQ, which is superimposed on the full sibling difference of 0.866, giving a standardized effect size/d of 0.008 / 0.866 = 0.0092 Let me try a power simulation: n <- 10000 siblings <- data.frame( sibling1AncestryPercentage = rnorm(n, mean=25, sd=1), sibling1NonancestryIQ = rnorm(n, mean=0, sd=12), sibling2AncestryPercentage = rnorm(n, mean=25, sd=1), sibling2NonancestryIQ = rnorm(n, mean=0, sd=12)) siblings$sibling1TotalIQ <- with(siblings, sibling1NonancestryIQ + sibling1AncestryPercentage*(0.008*15))
siblings$sibling2TotalIQ <- with(siblings, sibling2NonancestryIQ + sibling2AncestryPercentage*(0.008*15)) siblings$siblingAncestryDifference <- with(siblings, sibling1AncestryPercentage - sibling2AncestryPercentage)
siblings$siblingIQDifference <- with(siblings, sibling1TotalIQ - sibling2TotalIQ ) summary(siblings) # ... # siblingAncestryDifference siblingIQDifference # Min. :-5.370128122 Min. :-68.2971343 # 1st Qu.:-0.932086950 1st Qu.:-11.7903864 # Median : 0.002384529 Median : -0.2501536 # Mean : 0.007831583 Mean : -0.4166863 # 3rd Qu.: 0.938513265 3rd Qu.: 11.0720667 # Max. : 5.271052675 Max. : 67.5569825 summary(lm(siblingIQDifference ~ siblingAncestryDifference, data=siblings)) # ...Coefficients: # Estimate Std. Error t value Pr(>|t|) # (Intercept) -0.4192761 0.1705125 -2.45892 0.0139525 # siblingAncestryDifference 0.3306871 0.1220813 2.70874 0.0067653 # # Residual standard error: 17.05098 on 9998 degrees of freedom # Multiple R-squared: 0.000733338, Adjusted R-squared: 0.0006333913 # F-statistic: 7.337294 on 1 and 9998 DF, p-value: 0.006765343 confint(lm(siblingIQDifference ~ siblingAncestryDifference, data=siblings)) # 2.5 % 97.5 % # (Intercept) -0.75351500523 -0.08503724643 # siblingAncestryDifference 0.09138308561 0.56999105507 admixtureTest <- function(n, alpha=0.05, ancestryEffect=0.008) { siblings <- data.frame( sibling1AncestryPercentage =pmax(0, rnorm(n, mean=25, sd=1)), sibling1NonancestryIQ = rnorm(n, mean=0, sd=12), sibling2AncestryPercentage = pmax(0,rnorm(n, mean=25, sd=1)), sibling2NonancestryIQ = rnorm(n, mean=0, sd=12)) siblings$sibling1TotalIQ <- with(siblings, sibling1NonancestryIQ + sibling1AncestryPercentage*(ancestryEffect*15))
siblings$sibling2TotalIQ <- with(siblings, sibling2NonancestryIQ + sibling2AncestryPercentage*(ancestryEffect*15)) siblings$siblingAncestryDifference <- with(siblings, sibling1AncestryPercentage - sibling2AncestryPercentage)
siblings$siblingIQDifference <- with(siblings, sibling1TotalIQ - sibling2TotalIQ ) p <- summary(lm(siblingIQDifference ~ siblingAncestryDifference, data=siblings))$coefficients[8]
return(p<alpha)
}

power <- function(n, iters=10000, n.parallel=8) {
library(parallel)
library(plyr)
mean(unlist(mclapply(1:n.parallel, function(i) {

# powers <- sapply(seq(100, 10000, by=10), power)
power(100)
# [1] 0.0502
power(500)
# [1] 0.0535
power(1500)
# [1] 0.0642
power(15000)
# [1] 0.2251
power(50000)
# [1] 0.6077

So the estimated sample size is extremely large, well into the scores of thousands. This is large enough that it will be some time before biobanks or population samples are well-powered: participants may not have siblings, those siblings may be included, only ~15% of the American population is AA, all participants must be sequenced, and so on. If it requires n=70,000, half of participants have a matching sibling, and it’s drawn proportionally from the general population, that would imply that a total sample size of almost 1m. (For comparison, that’s about twice the size of the UK Biobank, and the US Precision Medicine Initiative cohort program aims for 1m total participants by 2020.)

Simplifications aside, it is difficult to see any way to bring this method down into the low thousands range, as that would require siblings to vary tremendously more in ancestry, have much more homogeneous IQs than they do, or ancestry to be vastly more potent than it could possibly be.

One possibility is to drop the idea of using only biological siblings. Is there any way to have ancestry differences as large as in the general population, but within a family? Half-siblings come to mind but those more typically tend to join the household as older kids or teenagers, so aren’t so good. One possibility is adoptees: there are a substantial number of African-American children adopted into other African-American households (white parents adopting black children is controversial and rarer, described as decreasing dramatically, but still substantial in total numbers, at least 20,000), and even a noticeable number of African children adopted abroad (14,800 from Ethiopia just 1999-2014, with more adoption from Nigeria & the Congo). The same logic of the within-family study should apply but to unrelated siblings who will have far greater differences in ancestry now (possibly anywhere up to 50% if an African child is adopted into an African-American family with considerable white ancestry & some luck). This would increase power dramatically, perhaps enough to bring the study within the realm of near-future feasibility.

Examining adoptions of African children would not be a convincing way of establishing group differences, particularly for IQ, as there are many known environmental harms (eg pre-natal lack of iodine is known to cause large impairments in cognitive which cannot be repaired later in life, and poor iodization is frequent in Africa), so while examining African adoptees would doubtless require a very small sample size, the results would be uninterpretable. So the more interesting case is instead examining AA adoptees/siblings, all of whom are raised in a wealthy (and iodine-sufficient) industrialized country.

In this case, we’re considering a pair of an AA sibling with the same IQ & ancestry distributions, as compared with adoptees who are either African (100% African ancestry) or likewise have the same IQ/ancestry distributions. Since the sibling & adoptee are unrelated, they effectively vary & differ as much as two random people from the general population would in IQ & African ancestry, except for shared-environment effects on IQ; shared-environment for adult IQ is relatively low, maybe 10% of variance. So instead of an SD of 15, they would vary moderately less, like 14.23 points (sqrt(15^2 * 0.9)).

One assumption here is a shared mean: one would actually expect, given the association of lighter skin with higher wealth/SES and darker with lower wealth/SES, that the adopting parents (and hence their biological children) would be relatively high on European ancestry, and conversely, the mothers giving up children for adoption would be relatively low, so the expected difference in ancestry is higher than simulated. Assuming equal means, however, is a conservative assumption since if such a correlation holds, the differences will be larger, hence the ancestry effect sizes larger, hence smaller sample sizes required. In the extreme version of this, the adoptive family is white and so the ancestry difference is maximal (~99% vs ~20%), requiring even smaller sample sizes, but at the cost of introducing complications like whether there are interactions with the white adoptive family not present in an AA adoptive family; in any case, such trans-racial adoption is apparently unpopular now, so it may not come up much.

Mean population European ancestry & population standard deviation

Ancestry effects remain as before; the mean ancestry is not too important as long as it’s not near 0, but since adoptees are drawn from the general population, the ancestry SD must be adjusted but it’s unclear what the right SD here is - cited studies range from 4% up to 11%, and this is a key parameter for power (with 4%, then sibling and adoptee will tend to be quite similar on ancestry percentage & much more data will be required, but with 11% they will differ a good deal and make results stronger).

Reported figures from the genetics literature for European ancestry in US African-American range from 14% to 24%, reflecting both sampling error and various biases & self-selection & geographic/regional effects in the datasets:

Considering just studies with usable ancestry percentages, population SD, and n, and using inferred SDs from Signorello:

admixture <- read.csv(stdin(), header=TRUE, colClasses=c("factor", "numeric", "numeric", "integer"))
Study,Mean,SD,N
"Halder et al 2008",0.143,0.133,136
"Ducci et al 2009",0.07,0.09,864
"Signorello et al 2010",0.071,0.08,379
"Bryc et al 2010",0.185,0.0465,365
"Nassir et al 2012",0.225,0.147,11712
"Bryc et al 2014",0.24,0.17,5269
"Kirkegaard et al 2016",0.17,0.11,140

# what is the standard error/precision of a population SD? http://davidmlane.com/hyperstat/A19196.html
admixture$SD.SE <- (0.71*admixture$SD) / sqrt(admixture$N) summary(admixture) # Study Mean SD N SD.SE # Bryc et al 2010 :1 Min. :0.0700000 Min. :0.0465000 Min. : 136.0 Min. :0.0009644066 # Bryc et al 2014 :1 1st Qu.:0.1070000 1st Qu.:0.0850000 1st Qu.: 252.5 1st Qu.:0.0016954481 # Ducci et al 2009 :1 Median :0.1700000 Median :0.1100000 Median : 379.0 Median :0.0021739221 # Halder et al 2008 :1 Mean :0.1577143 Mean :0.1109286 Mean : 2695.0 Mean :0.0034492579 # Kirkegaard et al 2016:1 3rd Qu.:0.2050000 3rd Qu.:0.1400000 3rd Qu.: 3066.5 3rd Qu.:0.0047591374 # Nassir et al 2012 :1 Max. :0.2400000 Max. :0.1700000 Max. :11712.0 Max. :0.0080973057 # Signorello et al 2010:1 library(metafor) r.mean <- rma(yi=Mean, sei=SD/sqrt(N), measure="SMD", ni=N, data=admixture); r.mean # Random-Effects Model (k = 7; tau^2 estimator: REML) # # tau^2 (estimated amount of total heterogeneity): 0.0046 (SE = 0.0027) # tau (square root of estimated tau^2 value): 0.0680 # I^2 (total heterogeneity / total variability): 99.82% # H^2 (total variability / sampling variability): 566.51 # # Test for Heterogeneity: # Q(df = 6) = 3477.2614, p-val < .0001 # # Model Results: # # estimate se zval pval ci.lb ci.ub # 0.1578 0.0258 6.1187 <.0001 0.1072 0.2083 predict(r.mean) # pred se ci.lb ci.ub cr.lb cr.ub # 0.1578 0.0258 0.1072 0.2083 0.0153 0.3003 r.sd <- rma(yi=SD, sei=SD.SE, measure="SMD", ni=N, data=admixture); r.sd # Random-Effects Model (k = 7; tau^2 estimator: REML) # # tau^2 (estimated amount of total heterogeneity): 0.0018 (SE = 0.0011) # tau (square root of estimated tau^2 value): 0.0425 # I^2 (total heterogeneity / total variability): 99.77% # H^2 (total variability / sampling variability): 440.67 # # Test for Heterogeneity: # Q(df = 6) = 3819.2793, p-val < .0001 # # Model Results: # # estimate se zval pval ci.lb ci.ub # 0.1108 0.0162 6.8587 <.0001 0.0792 0.1425 predict(r.sd) # pred se ci.lb ci.ub cr.lb cr.ub # 0.1108 0.0162 0.0792 0.1425 0.0216 0.2001 par(mfrow=c(2,1)) forest(r.mean, slab=admixture$Study)
forest(r.sd, slab=admixture$Study) There is high heterogeneity, as expected, and the meta-analytic summary are consistent with simply taking the mean, so meta-analysis was not really necessary. The issue of heterogeneity depends on how one wants to interpret these numbers: as the true latent African-American population mean/SD of European ancestry, or as a way to estimate the possible spread of sampling? In the former, the heterogeneity is a serious issue because it suggests the estimate may be badly biased or at least is highly imprecise; in the latter, it is both a curse and a benefit, since it implies that it is possible to recruit for genetics studies samples with a wide range of ancestry (thereby greatly increasing statistical power) but also that one might get unlucky & wind up with a very ancestry-homogeneous sample (if the sample turns out to have an SD as high as 20%, excellent; if it’s as low as 7.9%, one is in trouble). So for power analysis one might check the meta-analytic mean case, as well as the prediction interval (a 95% CI around the SD/mean does not mean that 95% of the true effects, including the inherent heterogeneity, will fall in that interval): SDs of 2%, 11%, and 20%. (For any cost-benefit analysis or trying to optimize expenditures, one would want to work with the posterior distributions to average over everything, but for just general informative purposes, those 3 are good parameters to check.) Power simulation Code: adopteeTest <- function(n, alpha=0.05, ancestryEffect=0.008, populationAncestryMean=0.1440, populationAncestrySD=0.1008, c=0.1) { unrelatedSiblingSD <- sqrt(15^2 * (1-c)) # subtract 10% for same shared-environment siblings <- data.frame( sibling1AncestryPercentage = pmax(0, rnorm(n, mean=populationAncestryMean*100, sd=populationAncestrySD*100)), sibling1NonancestryIQ = rnorm(n, mean=0, sd=unrelatedSiblingSD), adopteeAncestryPercentage = pmax(0, rnorm(n, mean=populationAncestryMean*100, sd=populationAncestrySD*100)), adopteeNonancestryIQ = rnorm(n, mean=0, sd=unrelatedSiblingSD)) siblings$sibling1TotalIQ <- with(siblings, sibling1NonancestryIQ + sibling1AncestryPercentage*(ancestryEffect*15))
siblings$adopteeTotalIQ <- with(siblings, adopteeNonancestryIQ + adopteeAncestryPercentage*(ancestryEffect*15)) siblings$siblingAncestryDifference <- with(siblings, sibling1AncestryPercentage - adopteeAncestryPercentage)
siblings$siblingIQDifference <- with(siblings, sibling1TotalIQ - adopteeTotalIQ ) p <- summary(lm(siblingIQDifference ~ siblingAncestryDifference, data=siblings))$coefficients[8]
return(p<alpha)
}

power <- function(n, sd, iters=10000, n.parallel=8) {
library(parallel)
library(plyr)
mean(unlist(mclapply(1:n.parallel, function(i) {

ns <- seq(100, 10000, by=100)
powerLow  <- sapply(ns, function(n) { power(n, sd=0.0216)})
powerMean <- sapply(ns, function(n) { power(n, sd=0.1108)})
powerHigh <- sapply(ns, function(n) { power(n, sd=0.2001)})

library(ggplot2); library(gridExtra)
pl <- qplot(ns, powerLow)  + coord_cartesian(ylim = c(0,1))
pm <- qplot(ns, powerMean) + coord_cartesian(ylim = c(0,1))
ph <- qplot(ns, powerHigh) + coord_cartesian(ylim = c(0,1))
grid.arrange(pl, pm, ph, ncol=1)

So for the worst-case SD, sample size is unclear but n>20,000 pairs; mean SD, n=1100 pairs; high SD, n=500 pairs. The latter two are feasible amounts for population registries or adoption-focused cohort studies. Thus genome adoption studies, combined with the much less powerful but more common within-sibling comparisons, are capable of delivering precise answers to long-standing questions about the origins of group differences with moderate sample sizes.

Operating on an aneurysm

In the excellent neurosurgery memoir Do No Harm: Stories of Life, Death, and Brain Surgery (Henry Marsh 2014), chapter 2 Aneurysm, there is a passage on weighing the costs of action and inaction:

A thirty-two-year-old woman, he said tersely. For surgery today. Had some headaches and had a brain scan. As he talked a brain scan flashed up on the wall.

It’s an unruptured aneurysm, seven millimetres in size, Fiona - the most experienced of the registrars - said. So there’s a point zero five per cent risk of rupture per year according to the international study published in 1998. And if it ruptures? Fifteen per cent of people die immediately and another thirty per cent die within the next few weeks, usually from a further bleed and then there’s a compound interest rate of four per cent per year.

…If we did nothing the patient might eventually suffer a haemorrhage which would probably cause a catastrophic stroke or kill her. But then she might die years away from something else without the aneurysm ever having burst. She was perfectly well at the moment, the headaches for which she had had the scan were irrelevant and had got better. The aneurysm had been discovered by chance. If I operated I could cause a stroke and wreck her - the risk of that would probably be about four or five per cent. So the acute risk of operating was roughly similar to the life-time risk of doing nothing. Yet if we did nothing she would have to live with the knowledge that the aneurysm was sitting there in her brain and might kill her any moment.

Reading this, I was a little surprised by Marsh’s evaluation given those specific numbers. Intuitively, it did not seem to me that a single risk of 5% was anywhere near as bad as a lifelong risk of 0.5%, for a 32 year old woman who would probably live another 50 years - the one number is 10x bigger than the other, but the other number is 50x bigger, and a quick heuristic for the total probability of many independent small probabilities is to just sum them up, suggesting that the risk of the untreated aneurysm was much worse (50*0.005=0.25, and 0.25>0.05). So I thought after I finished reading the book, I would work it out a little more accurately.

Risk

Specifically, this is a 32yo woman and the UK female life expectancy is ~80yo in 2015, so she had ~48 years left. The consequences of the aneurysm bursting is a large chance of instant death or else severe disability with death to soon follow; the consequence of surgery going wrong is also instant death or severe disability, presumably with a high chance of death soon following, so it looks like we can assume that the bad outcome in either case is the same. what is the probability of the aneurysm never bursting in all 48 years? (1-0.005)^48 = 0.786, or a probability of bursting of 21%. 21% is 4x larger than 5%. Since 21% is 4x larger and the consequences are similar, this would suggest that the risks are not roughly similar and it looks much worse to not operate.

Expected loss

But that’s just the risk of an event, not the expected loss:

1. In the case of doing surgery immediately, the expected loss, with years treated equally and a 5% instant risk from operation, is simply 48 * 0.005 = 0.24 years of life; all 48 years are risked on a single throw of the surgical dice, but after that she is safe.
2. In the case of doing nothing and letting the aneurysm stay with a 0.5% annual risk from non-operation, it’s not as simple as 48 * 0.21 = 10.1 years, because you cannot die of an aneurysm if you died in a previous year. The risk will instead follow a negative binomial distribution (number of years until 1 failure), and then the loss is the 48 years minus however many she actually got. That’s not the same as the expectation of the negative binomial, which in this case is 200 years (the expectation of a negative binomial with 1 failure and a success rate of 1-0.005 is 1/(1-(1-0.005))=200) and she will die of other causes before then, in which case the aneurysm turned out to be harmless.

We can simulate many draws from the negative binomial, ignore as 0 any time where the aneurysm struck after her life expectancy of 48 more years is past, hold onto the losses, and calculate the mean loss: mean(sapply(rnbinom(10e4, 1, 0.005), function(years) { if(years>48) { 0; } else { 48-years; }})) ~> 5.43.

So the expected loss from surgery looks even better than the risk did, as it is 22.6x smaller.

What about adjusting for older years being less valuable? We might say that the surgery look unfairly good because we are ignoring how its losses are front-loaded in the 30s, some of the best years of one’s life, and treating a loss of her 33rd year as being as bad as a loss of her 48th year. In terms of age weighting, DALYs usually use a 3% annual discounting; DALYs and QALYs differ in some ways but for this analysis I think we can treat them as equivalent and use the DALY age-discounting to calculate our QALYs. So we can redo the two expected losses including the discounting to get:

1. Surgery: 0.05 * sum((1-0.03)^(0:48)) ~> 1.291
2. No surgery: mean(unlist(sapply(sapply(rnbinom(10e4, 1, 0.005), function(years) { if(years>48) { 0; } else { 48-years; }}), function(yr) { sum((1-0.03)^(0:yr)); }))) ~> 4.415

By appropriately penalizing the surgery’s loss of high-quality early years as compared to the aneurysm’s loss of just some elderly years, the surgery’s superiority falls to 3.4x, and the gain is 3.124. (And if we include the mental wellbeing of the woman as a final touch, the surgery looks even better.)

How sensitive is the surgical superiority to the parameters?

• Surgical risk: a 4x increase in risk to 20% would create parity
• Aneurysm risk: if the annual risk of aneurysm were as low as 0.04% rather than 0.5%, then there would be parity
• Life expectancy & discount rate: no change will reverse the ordering

It seems extremely unlikely that Marsh could be as wrong about the surgical risk as to mistake 5% for 20%, especially for an operation he says he used to do routinely, and it also seems unlikely that the study on the annual risk of an aneurysm bursting could be as far off as 10x, so the difference is solid.

Finally, having a surgery is much more expensive than not having it. Surgery is always expensive, and neurosurgery undoubtedly so - elsewhere in the book, Marsh quotes an American neurosurgeon’s estimate of $100,000 for a particularly complex case. Clipping an aneurysm surely cannot cost that much (being both much simpler and also being done in a more efficient healthcare system), but it’s still not going to be trivial. Does the cost of aneurysm surgery outweigh the benefit? To convert the DALY loss to a dollar loss, we could note that UK PPP per capita is ~$38,160 (2013) so the gain from surgery would be (4.415 - 1.291) * 38169=$119k, well above the$100k worst-case. Or more directly, the UK NHS prefers to pay <£20,000 per QALY and will generally reject treatments which cost >£30,000 per QALY as of 20072 (implying QALYs are worth somewhat less than £30,000); the median US 2008 hospital cost for clipping an aneurysm is $36,188 or ~£23,500; and the gain is 3.124 QALYs for ~£7500/QALY - so clipping the aneurysm in this case definitely clears the cost-benefit threshold (as we could have guessed from the fact that in the anecdote, the NHS allows her to have the surgery). After calculating the loss of years, differing values of years, and cost of surgery, the surgery still comes out as substantially better than not operating. The Power of Twins: Revisiting Student’s Scottish Milk Experiment Example Randomized experiments require more subjects the more variable each datapoint is to overcome the noise which obscures any effects of the intervention. Reducing noise enables better inferences with the same data, or less data to be collected, which can be done by balancing observed characteristics between control and experimental datapoints. A particularly dramatic example of this approach is running experiments on identical twins rather than regular people, because twins vary far less from each other than random people do. In 1931, the great statistician Student noted problems with an extremely large (n=20,000) Scottish experiment in feeding children milk (to see if they grew more in height or weight), and claimed that the experiment could have been done far more cost-effectively with an extraordinary reduction of >95% fewer children if it had been conducted using twins. He, however, did not provide any calculations or data demonstrating this. I revisit the issue and run a power calculation on height indicating that Student’s claims were correct and that the experiment would have required ~97% fewer children if run with twins. This reduction is not unique to the Scottish experiment and in general, one can expect a reduction of 89% using twins rather than regular people. Simple randomization vs blocking Randomized experiments enable causal inference by assuring that the experimental group is identical, on average, in all ways to the control group and the subsequent differences are caused by the intervention. If a coin is flipped, then each group will have the same fraction of women, the same fraction of atheists, the same fraction of people with a particular SNP on COMT, the same fraction of people with latent cancers, etc - on average. But since there are thousands upon thousands of ways in which subjects can differ which might affect the results (just consider how almost all human traits are genetically heritable), and there may only be a few dozen subjects, randomization cannot guarantee exact or even approximate similarity of groups on every single trait and with same samples can generate grossly imbalanced samples (perhaps you have 2 groups of 10, and one group turns out to have 9 women and the other just 1). This doesn’t bias the results or defeat the point of randomization, but it can add a lot of noise, reducing our efficiency, thereby making our existing studies less meaningful, requiring more expensive studies to estimate effects to a useful level of precision, and blocking profitable decisions. When put this way, one might wonder how to improve the design of simple randomized experiments: if you have a group with too many women and too few men, why not change your coin flip to instead pairs of women and men? Instead of flipping a fair coin for each patient to pick whether that patient goes into the control or experimental group, why not instead take a pair of women and flip a coin to decide whether the one on the left goes into the experimental group, and then the woman on the right goes into the other group? And likewise for the men. In fact, why not try to extend this matching up to as many things as you can measure? If we use identical twins (like in Gustav III of Sweden’s coffee experiment), we match them on not just nationality, gender, location, parents, age, food consumption or whatnot, we even match them on genetics too, which is part of why identical twins are so eerily identical even when raised apart. (Note that this point isn’t the same as the variance component estimating going on in twin study or other family designs; we care that they are very similar and thus comparisons are less noisy, not the question of how much of that eerie similarity is due to genetics or other sources.) Indeed, since a person is matched with themselves on just about everything and are even better than identical twins, why not test the treatment on the same person over multiple time periods? But people do change over time and there might be lingering effects, so perhaps it would be even better if you can test it on the same person simultaneously: for example, test an acne cream A on one side of the face and acne cream B on the other half of each subject’s face, flipping a coin to decide whether A/B or B/A. This is the concept of blocking, one of whose advocates was Student, the great early statistician. Efficiency of blocking for the Lanarkshire Milk Experiment The history of blocking and its advantages compared to simple randomized experiments are discussed at length in Balanced versus Randomized Field Experiments in Economics: Why W. S. Gosset aka Student Matters, Ziliak 2014. Ziliak writes (pg22) about an interesting example where Student calculates that a well-blocked sample of 100 could be better than 20,000 with simple randomization: The intuition behind the higher power of ABBA and other balanced designs to detect a large and real treatment difference was given by Student in 1911.68 Now if we are comparing two varieties it is clearly of advantage to arrange the plots in such a way that the yields of both varieties shall be affected as far as possible by the same causes to as nearly as possible an equal extent.69 He used this principle of maximum contiguity often, for example when he for example when he illustrated the higher precision and lower costs that would be associated with a small-sample study of biological twins, to determine the growth trajectory of children fed with pasteurized milk, unpasteurized milk, and no milk at all, in The Lanarkshire Milk Experiment (Student, 1931a).69 Student (1931a, p. 405) estimated that 50 pairs of [identical twins] would give more reliable results than the 20,000 child sample, neither balanced nor random, actually studied in the experiment funded by the Scotland Department of Health. [I]t would be possible to obtain much greater certainty in the measured difference of growth in height and weight of children drinking raw versus pasteurized milk at an expenditure of perhaps 1-2% of the money and less than 5% of the trouble. Likewise, Karlan and List (2007, p. 1777) could have revealed more about the economics of charitable giving - for less - using a variant of Student’s method. Instead the AER article studied n = 50,083 primarily white, male, pro-Al Gore donors to public radio, neither random nor balanced. I thought this was an awesome example and topical inasmuch as raw milk is still an issue because gourmands prize it over pasteurized milk (leading to occasional illnesses or deaths and subsequent legal/political maneuvering to ban/preserve access to raw milk), and I was curious about the details of how Student computed that the surprisingly low number of 100 twins (50 pairs) would suffice. So I looked for Student’s original paper. Student begins by summarizing the Lanarkshire experiment in a little more detail: In the spring of 1930, a nutritional experiment on a very large scale was carried out in the schools of Lanarkshire. For four months 10,000 school children received 0.75 pint of milk per day, 5000 of these got raw milk and 5,000 pasteurized milk, in both cases Grade A (Tuberculin tested3); another 10,000 children were selected as controls and the whole 20,000 children were weighed and their height was measured at the beginning and end of the experiment…The 20,000 children were chosen in 67 schools, not more than 400 nor less than 200 being chosen in any one school, and of these half were assigned as feeders and half as controls, some schools were provided with raw milk and the others with pasteurized milk, no school getting both…Secondly, the selection of the children was left to the Head Teacher of the school and was made on the principle that both controls and feeders should be representative of the average children between 5 and 12 years of age: the actual method of selection being important I quote from Drs Leighton and McKinlay’s [1930] Report: The teachers selected the two classes of pupils, those getting milk and those acting as controls, in two different ways. In certain cases they selected them by ballot and in others on an alphabetical system. Unfortunately, while ambitious, the randomization was heavily compromised: after invoking the goddess of chance they unfortunately wavered in their adherence to her for we read: In any particular school where there was any group to which these methods had given an undue proportion of well fed or ill nourished children, others were substituted in order to obtain a more level selection. This is just the sort of afterthought that most of us have now and again and which is apt to spoil the best laid plans. In this case it was a fatal mistake, for in consequence the controls were, as pointed out in the Report, definitely superior both in weight and height to the feeders by an amount equivalent to about 3 months’ growth in weight and 4 months’ growth in height. Presumably this discrimination in height and weight was not made deliberately, but it would seem probable that the teachers, swayed by the very human feeling that the poorer children needed the milk more than the comparatively well to do, must have unconsciously made too large a substitution of the ill-nourished among the feeders and too few among the controls and that this unconscious selection affected, secondarily, both measurements. Student then observes that besides producing a baseline imbalance (which is clearly visible in the graphs on pg400 & pg402, where the controls are taller in every time period, although this advantage lessens with time), this favoring of the poor could produce a systematic bias in the recorded weights during winter, which were made with the childrens’ clothes on, as it is entirely possible that poorer children will have lighter or fewer heavy winter clothes. Another issue is allocating entire schools to using either pasteurized or raw milk as their comprison to the no-milk controls (leading to problems in accounting for the hierarchical nature of the data due to confounding of school level effects with the pasteurized/raw effect). These three issues are reflected in anomalies in the data, reducing our confidence in the results despite it having been a randomized experiment (rather than something lamer like a survey noting that children who could afford milk were taller). Having performed the post-mortem on the Lanarkshire Milk Experiment, Student then proposes improvements to the design, culminating in the most radical (pg405): (2) If it be agreed that milk is an advantageous addition to children’s diet - and I doubt whether any one will combat that view - and that the difference between raw and pasteurised milk is the matter to be investigated, it would be possible to obtain much greater certainty at an expenditure of perhaps 1-2% of the money [This is a serious consideration: the Lanarkshire experiment cost about £7500. [Which is ~£374,800 in 2016 pounds sterling or ~$541,100.]] and less than 5% of the trouble.

For among 20,000 children there will be numerous pairs of twins; exactly how many it is not easy to say owing to the differential death rate, but, since there is about one pair of twins in 90 births, one might hope to get at least 160 pairs in 20,000 children. But as a matter of fact the 20,000 children were not all the Lanarkshire schools population, and I feel pretty certain that some 200-300 pairs of twins would be available for the purpose of the experiment. Of 200 pairs some 50 would be identicals and of course of the same sex, while half the remainder would be non-identical twins of the same sex.

Now identical twins are probably better experimental material than is available for feeding experiments carried out on any other mammals, and the error of the comparison between them may be relied upon to be so small that 50 pairs of these would give more reliable results than the 20,000 with which we have been dealing. The proposal is then to experiment on all pairs of twins of the same sex available, noting whether each pair is so similar that they are probably identicals or whether they are dissimilar. Feed one of each pair on raw and the other on pasteurised milk, deciding in each case which is to take raw milk by the toss of a coin. Take weekly measurements and weigh without clothes. Some way of distinguishing the children from each other is necessary or the mischievous ones will play tricks. The obvious method is to take finger-prints, but as this is identified with crime in some people’s minds, it may be necessary to make a different indelible mark on a fingernail of each, which will grow off after the experiment is over. With such comparatively small numbers further information about the dietetic habits and social position of the children could be collected and would doubtless prove invaluable. The comparative variation in the effect in identical twins and in unlike twins should furnish useful information on the relative importance of Nature and Nurture. …[The twin experiment] is likely to provide a much more accurate determination of the point at issue, owing to the possibility of balancing both nature and nurture in the material of the experiment.

This is a reasonable suggestion, but I was disappointed to see that Student was, it seems, effectively guessing, as he gives no calculation or reference to another work with a similar calculation. (If Ziliak was going to cite it as an example, it would’ve been better if he had verified that Student’s speculation was right at least to within an order of magnitude rather than simply quoting him as an authority.)

Power estimate of twins vs general population

The calculation doesn’t seem hard. For this example, using height, it’d go something like:

1. take the estimated height gain in inches
2. find the distribution of height differences for twins and find the distribution of height for the general population of Scottish kids those ages
3. convert the inch gain into standard deviations for twins and for general population
4. plug those two into an effect size calculator asking for, say, 80% power
5. compare how big n1 you need of twin pairs and how many n2 of general population, and report the fraction $\frac{2\cdot n_1}{n_2}$ and how close it is to 5%.

Milk’s effect on male height

To start, on pg403, a table reports Gain in height in inches by Feeders over Controls, for which the largest effect in boys is the 5-7yo boys group at +0.083(.011) inches. So we are targeting an average increase of a tenth of an inch.

What is the height variability or standard deviations for twins and Lanarkshire children of a similar age? The followup paper The Lanarkshire Milk Experiment, Elderton 1933 helpfully reports standard deviations both for Lanarkshire children and cites some standard deviation for twins’ heights at that time (pg2):

Dr Stocks in his study of twins [Percy Stocks 1930, assisted by Mary N. Karn: A Biometric Investigation of Twins and their Brothers and Sisters, Annals of Eugenics, Vol. v, pp. 46-50. Francis Galton Laboratory for National Eugenics.] found differences in weight as great as 28 hectograms (10 ounces) in those twins he regarded as monozygotic whose ages corresponded to the children in the milk experiment. The standard deviation of weight in pounds is roughly twice that of the standard deviation of height in inches, so that if 8 ounces difference in initial weight be permitted, 1/4 inch difference in height could be allowed. Judging also by Dr Stocks’ material in which monozygotic twins showed a modal difference of 1 cm in height it would have been justifiable to allow children to be paired who differed by two-eighths of an inch, but the labour of pairing would have been much heavier if a greater variation than that entered on the cards had been allowed for height as well as for weight…In Table I the standard deviations and coefficients of variation of the initial height and weight for each year of birth are given, and if these be compared with those for Glasgow boys and girls, it will be seen that they are distinctly less. The Glasgow figures were obtained by linear interpolation and are given in brackets after those for the selected Lanarkshire data.

Elderton’s Table 1 reports on the Lanarkshire children’s grouped data by 6 years 9 months, 7 years 9 months, and then 8 years 9 months & higher; the first two presumably map best onto our 5-7yo group, and have values of n=382 with standard deviation 1.483(2.58), and n=337 with standard deviation 1.648(2.82); pooling the standard deviations, we get a standard deviation of 1.56 for the general population.

Elderton’s summary of Stocks’s twin research is confusingly worded (partly because Stocks worked in centimeters and Elderton inches), but she appears to be saying that the standard deviation of differences in Stocks’s twins is 0.25 inches, which compared with 1.483 is much smaller and around one-fifth; Table II (pg11) in Stocks 1930 records variability of identical twins vs fraternal vs their non-twin siblings in mean corrected Deviates, mentioning that the root-mean-squared difference in the table is the standard deviation, so the σ0 of height for identical twins is 0.9497 while the general population is 6.01cm (pg13), and $\frac{0.9497}{6.01}$ comes out to one-fifth, confirming where Elderton got her specific estimate of ~0.25 inches as the standard deviation for identical twins.

So the claimed effect is +0.083 inches, which represents d=0.05 (for the general population) and d=0.332 (twin differences). We then ask how much data is required to conduct a well-powered experiment to detect the existence of such an effect:

generalD <- 0.083/1.560322176
twinD <- 0.083/0.25
generalP <- power.t.test(d=generalD, power=0.8); generalP
#      Two-sample t test power calculation
#
#               n = 5548.623219
#           delta = 0.05319414239
#              sd = 1
#       sig.level = 0.05
#           power = 0.8
#     alternative = two.sided
#
# NOTE: n is number in *each* group
twinP <- power.t.test(d=twinD, power=0.8); twinP
#      Two-sample t test power calculation
#
#               n = 143.3836014
#           delta = 0.332
#              sd = 1
#       sig.level = 0.05
#           power = 0.8
#     alternative = two.sided
#
# NOTE: n is number in *each* group
twinP$n / generalP$n
# [1] 0.02584129355

So with 143 twin-pairs or n=286, we can match the power of a sample drawn from the general population with 5548 in each group or n=11096 - savings of ~98% of the sample. (To put that in perspective, if costs scaled exactly per head, then that estimated cost of ~$541,100 would have instead been$13,982, for a savings of $527,117.) Student’s guess of 1-2% proves to be on the money, and the experiment is also feasible as 143 twin-pairs is well below the number of twin-pairs that Student estimated to be available (>160, and more likely 200-300; 300 twin-pairs would yield a power of ~98%, exceeding the Lanarkshire’s 20000’s <96% power). We can safely say that Student’s Scottish milk experiment example does indeed demonstrate the power of twins. All traits We can go further and estimate the power of twins in general for experimentation. While height is somewhat unusually heritable, we can say with confidence that almost all traits studied are highly heritable based on the previously mentioned mega-meta-analysis Meta-analysis of the heritability of human traits based on fifty years of twin studies, Polderman et al 2015 which compiles data on 17,804 traits estimated from ~14.5m pairs. The upshot is that averaging over all those traits, 48.8% of variance is due to heritability and 17.4% shared-environment4; implying that compared to a sample drawn from the general population (who share neither genetics nor the shared-environment upbringing), identical twins will have 33.8% (1-(0.488+0.174)) of the variability. For easier comparison with the running example, we can redo the height calculation but assuming we are looking at a generic trait with a higher variability twinAll <- 0.083/(0.338*1.560322176) generalAll <- 0.083/1.560322176 generalAll <- power.t.test(d=generalD, power=0.8) twinAll <- power.t.test(d=twinAll, power=0.8) twinAll$n / generalAll$n # [1] 0.1143975625 In general, an experiment run using twins will require a sample 11% the size of the same experiment run using the general population. RNN metadata for mimicking individual author style Char-RNNs are unsupervised generative models which learn to mimic text sequences. I suggest extending char-RNNs with inline metadata such as genre or author prefixed to each line of input, allowing for better & more efficient metadata, and more controllable sampling of generated output by feeding in desired metadata. An experiment using torch-rnn on a set of ~30 Project Gutenberg e-books (1 per author) to train a large char-RNN shows that a char-RNN can learn to remember metadata such as authors, learn associated prose styles, and often generate text visibly similar to that of a specified author. Due to length, this has been split out to another page. MCTS An implementation in R of a simple Monte Carlo tree search algorithm (using Thompson sampling rather than a UCT) implemented with data.tree. This MCTS assumes binary win/loss (1/0) terminal rewards with no intermediate rewards/costs so it cannot be used to solve general MDPs, and does not expand leaf nodes in the move tree passed to it. (I also suspect parts of it are implemented wrong though it reaches the right answer in a simple Blockworld problem and seems OK in a Tic-Tac-Toe problem.) library(data.tree) ## MCTS helper functions: playOutMoves <- function(move, state, actions) { for (i in 1:length(actions)) { state <- move(state, actions[i])$State
}
return(state)
}
playOutRandom <- function(move, state, actions, timeout=1000, verbose=FALSE) {
action <- sample(actions, 1)
turn <- move(state, action)
if(verbose) { print(turn); };
if (turn$End || timeout==0) { return(turn$Reward) } else {
playOutRandom(move, turn$State, actions, timeout=timeout-1, verbose) } } createTree <- function(plys, move, moves, initialState, tree=NULL) { if (is.null(tree)) { tree <- Node$new("MCTS", win=0, loss=0) }
if (plys != 0) {
for(i in 1:length(moves)) {
x <- tree$AddChild(moves[i], win=0, loss=0) createTree(plys-1, move, moves, initialState, tree=x) } } # cache the state at each leaf node so we don't have to recompute each move as we later walk the tree to do a rollout tree$Do(function(node) { p <- node$path; node$state <- playOutMoves(move, initialState, p[2:length(p)]); }, filterFun = isLeaf)
return(tree)
}

mcts <- function (tree, randomSimulation, rollouts=1000) {
replicate(rollouts, {
# Update posterior sample for each node based on current statistics and use Thompson sampling.
# With a beta uniform prior (Beta(1,1)), update on binomial (win/loss) is conjugate with simple closed form posterior: Beta(1+win, 1+n-win).
# So we sample directly from that posterior distribution for Thompson sampling
tree$Do(function(node) { node$Thompson <- rbeta(1, 1+node$win, 1+(node$win+node$loss)-node$win) })
# find & run 1 sample:
node <- treeWalk(tree)
rollout <- randomSimulation(node$state) if(rollout==1) { node$win <- node$win+1; } else { node$loss <- node$loss+1; } # propagate the new leaf results back up tree towards root: tree$Do(function(x) { x$win <- Aggregate(x, "win", sum); x$loss <- Aggregate(x, "loss", sum) }, traversal = "post-order")
})
}

## walk the game tree by picking the branch with highest Thompson sample down to the leaves
## and return the leaf for a rollout
treeWalk <- function(node) {
if(length(node$children)==0) { return(node); } else { children <- node$children
best <- which.max(sapply(children, function(n) { n$Thompson; } )) treeWalk(children[[best]]) } } mctsDisplayTree <- function(tree) { tree$Do(function(node) { node$P <- node$win / (node$win + node$loss) } )
tree$Sort("P", decreasing=TRUE) print(tree, "win", "loss", "P", "Thompson") } ## Blockworld simulation ## 0=empty space, 1=agent, 2=block, 3=goal point blockActions <- c("up", "down", "left", "right") blockInitialState <- matrix(ncol=5, nrow=5, byrow=TRUE, data=c(0,0,0,0,1, 0,2,0,0,2, 0,0,0,2,0, 0,2,0,0,0, 0,0,0,0,3)) blockMove <- function(state, direction) { if(state[5,5] == 2) { return(list(State=state, Reward=1, End=TRUE)) } position <- which(state == 1, arr.ind=TRUE) row <- position[1]; col <- position[2] rowNew <- 0; colNew <- 0 switch(direction, # if we are at an edge, no change up = if(row == 1) { rowNew<-row; colNew<-col; } else { rowNew <- row-1; colNew <- col; }, down = if(row == 5) { rowNew<-row; colNew<-col; } else { rowNew <- row+1; colNew <- col; }, left = if(col == 1) { rowNew<-row; colNew<-col; } else { rowNew <- row; colNew <- col-1; }, right = if(col == 5) { rowNew<-row; colNew<-col; } else { rowNew <- row; colNew <- col+1; } ) # if there is not a block at the new position, make the move if (state[rowNew,colNew] != 2) { state[row,col] <- 0 state[rowNew,colNew] <- 1 return(list(State=state, Reward=0, End=FALSE)) } else { state[rowNew,colNew] <- 1 state[row,col] <- 0 switch(direction, # if the block is at the edge it can't move up = if(rowNew == 1) { } else { state[rowNew-1,colNew] <- 2 }, down = if(rowNew == 5) { } else { state[rowNew+1,colNew] <- 2 }, left = if(colNew == 1) { } else { state[rowNew,colNew-1] <- 2 }, right = if(colNew == 5) { } else { state[rowNew,colNew+1] <- 2 } ) # a block on the magic 5,5 point means a reward and reset of the playing field if(state[5,5] == 2) { return(list(State=state, Reward=1, End=TRUE)) } else { return(list(State=state, Reward=0, End=FALSE)) } } } ## Blockworld examples: # blockMove(blockInitialState, "left") # blockMove(blockInitialState, "down") # blockMove(blockInitialState, "right")$State
# blockMove(blockMove(blockInitialState, "right")$State, "down") # blockMove(blockMove(blockMove(blockInitialState, "down")$State, "down")$State, "down") # playOutMoves(blockMove, blockInitialState, c("down", "down", "down")) # playOutRandom(blockMove, blockInitialState, blockActions) tree <- createTree(2, blockMove, blockActions, blockInitialState) mcts(tree, function(state) { playOutRandom(blockMove, state, blockActions) }) mctsDisplayTree(tree) tree2 <- createTree(3, blockMove, blockActions, blockInitialState) mcts(tree2, function(state) { playOutRandom(blockMove, state, blockActions) }) mctsDisplayTree(tree2) ## Tic-Tac-Toe tttActions <- 1:9 tttInitialState <- matrix(ncol=3, nrow=3, byrow=TRUE, data=0) tttMove <- function(state, move) { move <- as.integer(move) # whose move is this? Player 1 moves first, so if the number of pieces are equal, it must be 1's turn: player <- 0; if(sum(state == 1) == sum(state == 2)) { player <- 1 } else { player <- 2} # check move is valid: if(state[move] == 0) { state[move] <- player } ## enumerate all possible end-states (rows, columns, diagonals): victory, or the board is full and it's a tie victory <- any(c( all(state[,1] == player), all(state[1,] == player), all(state[,2] == player), all(state[2,] == player), all(state[,3] == player), all(state[3,] == player), all(as.logical(c(state[1,1], state[2,2], state[3,3]) == player)), all(as.logical(c(state[1,3], state[2,3], state[3,1]) == player)) )) tie <- all(state != 0) # if someone has won and the winner is player 1, then a reward of 1 if(victory) { return(list(State=state, Reward=as.integer(player==1), End=TRUE)) } else { if(tie) { return(list(State=state, Reward=0, End=TRUE)) } else { return(list(State=state, Reward=0, End=FALSE)) } } } ## Tic-Tac-Toe examples: # tttMove(tttMove(tttMove(tttInitialState, 5)$State, 9)$State, 2) # playOutMoves(tttMove, tttInitialState, c(5, 9, 2)) # playOutRandom(tttMove, tttInitialState, tttActions, verbose=TRUE) treeTTT <- createTree(2, tttMove, tttActions, tttInitialState) mcts(treeTTT, function(state) { playOutRandom(tttMove, state, tttActions) }) mctsDisplayTree(treeTTT) ## hypothetical: if opponent plays center (5), what should be the reply? treeTTT2 <- createTree(2, tttMove, tttActions, tttMove(tttInitialState, 5)$State)
mcts(treeTTT2, function(state) { playOutRandom(tttMove, state, tttActions) })
mctsDisplayTree(treeTTT2)

Candy Japan A/B test

Due to length, has been split out to Candy Japan’s new box A/B test.

DeFries-Fulker power analysis

DeFries-Fulker (DF) extremes analysis

• DeFries & Fulker 1985
• DeFries et al 1987, Evidence for a genetic aetiology in reading disability of twins /docs/genetics/1987-defries.pdf
• DeFries & Fulker 1988, Multiple regression analysis of twin data: Etiology of deviant scores versus individual differences
• A Model-Fitting Implementation of the DeFries-Fulker Model for Selected Twin Data Purcell & Sham 2003 /docs/genetics/2003-purcell.pdf
• LaBuda et al 1986
• Gillespie & Neale 2006 A Finite Mixture Model for Genotype and Environment Interactions: Detecting Latent Population Heterogeneity http://citeseerx.ist.psu.edu/viewdoc/download?doi=10.1.1.534.6298&rep=rep1&type=pdf
• Purcell et al 2001 Comorbidity between verbal and non-verbal cognitive delays in 2-year-olds: A bivariate twin analysis https://www.researchgate.net/profile/Emily_Simonoff/publication/227715188_Comorbidity_between_verbal_and_nonverbal_cognitive_delays_in_2yearolds_a_bivariate_twin_analysis/links/0fcfd50b74917d5933000000.pdf
• http://www.sciencedirect.com/science/article/pii/S0160289614001676 Thinking positively: The genetics of high intelligence, Shakeshaft et al 2015

mixture model

generateSiblingPair <- function(ID=TRUE) {
## Population mean 100, SD 15; let's make family means distributed normally too;
## heritability 0.8, shared environment 0.1, siblings share half of genes on average + shared environment
## so a pair of siblings has 1 - (0.8*0.5+0.1) = 0.5 of the variance of the general population.
parental <- mean(rnorm(1,mean=100,sd=15*0.8), rnorm(1,mean=100,sd=15*0.8))
siblings <- rnorm(2, mean=parental, sd=15*(1 - (0.8*0.5+0.1)))
## Siblings will tend to vary this much, unless they are, lamentably, one of the, say,
## 5% struck by mutational lightning and reduced to an IQ of, let's say, 80
if(ID) { siblings <- ifelse(rbinom(2,1,prob=0.05), siblings,rnorm(2, mean=80, sd=15)) }
return(c(max(siblings), min(siblings)))
}
generateSiblingPairs <- function(n,ID=TRUE) { as.data.frame(t(replicate(n, generateSiblingPair(ID=ID)))) }
## dataset with lightning:
df <- round(rescale(generateSiblingPairs(1000000, ID=TRUE), mean=5, sd=2))
## floor/ceiling at 0/9 for everyone:
df[df$V1>9,]$V1 <- 9
df[df$V1<1,]$V1 <- 1
df[df$V2>9,]$V2 <- 9
df[df$V2<1,]$V2 <- 1

## dataset without:
df2 <- round(rescale(generateSiblingPairs(1000000, ID=FALSE), mean=5, sd=2))
df2[df2$V1>9,]$V1 <- 9
df2[df2$V1<1,]$V1 <- 1
df2[df2$V2>9,]$V2 <- 9
df2[df2$V2<1,]$V2 <- 1

par(mfrow=c(2,1))
hist(df$V1 - df$V2)
hist(df2$V1 - df2$V2)

## mixture modeling:
library(flexmix)
## check k=1 vs k=2 on df1, where k=2 is ground truth:
g1.1 <- flexmix(I(V1-V2) ~ 1, k=1, data=df)
g1.2 <- flexmix(I(V1-V2) ~ 1, k=2, data=df)
summary(g1.1); summary(g1.2)

## check k=1 vs k=2 on df2, where k=1 is ground truth:
g2.1 <- flexmix(I(V1-V2) ~ 1, k=1, data=df2)
g2.2 <- flexmix(I(V1-V2) ~ 1, k=2, data=df2)
summary(g2.1); summary(g2.2)

Inferring mean IQs from SMPY/TIP elite samples

Samples taken from the extremes of mixtures of distributions can have very different properties than random samples, such as wildly disproportionate representation of one distribution. This can be used to infer differing means. I demonstrate working backwards from the racial composition of TIP/SMPY samples of extremely (1-in-10,000) gifted youth to estimate the overall racial means, which is consistent with the known racial means and hence an unbiased selection process, using ABC to infer Bayesian credible intervals on the estimated means.

The general properties of statistical distributions can be very different from the properties of specific subsets in counterintuitive ways. One common example is that a small difference in means for many distributions can lead to large differences in extreme subsets.

For example, male and female average heights differ by a relatively small amount, inches at most. So in a random sample, plenty of women will be taller than men, and vice versa. However, if instead ask the sex of the tallest person in the sample, it will often be male, and the larger the sample, the more certain we can be that it will be male, and that the top X% by height will be male. Likewise, if we wanted to start a basketball league and recruited the tallest 100 people in the country, this small mean difference will show up as our entire basketball league turning out to be male. (And since height is highly heritable, we may find out that many of them are related!) What seemed like a small difference become a large one; we could have worked it out in advance if we had thought about it.

Reasoning from the general to the particular turned out to be tricky in this case because we were dealing with extreme values rather than random samples - 1 basketball player chosen by height from thousands of people. Many things of great interest turn out to be like that: we are interested in the extremes much more than the expectation. Running a 2-hour marathon is an extreme on athleticism; winning the Nobel is an extreme on scientific accomplishment; being enlisted in the NBA is an extreme on height; being admitted to MIT/Stanford/Harvard is an extreme on intelligence; murdering someone is an extreme on violence; winning an Academy Award is an extreme on acting success. When we ask questions like, why does the world record in this sport keep being shattered or why are so many NBA players related or how good can we expect the best chess player to be in 10 years or does this racial composition prove bias or how much more important are the best authors in literature than obscurer figures or why do so few women win the Field Medal, we’re asking extreme value questions whose answers may be counterintuitive - and the answer may be as simple as the shape of distributions, and a slightly lower mean here or a slightly higher standard deviation there. (Working backwards from a sample selected for passing a threshold to a mean can be called the method of limits or the method of thresholds.)

The study When Lightning Strikes Twice: Profoundly Gifted, Profoundly Accomplished, Makel et al 2016 describes the accomplishments of the Duke TIP sample, 259 children selected for their intelligence by taking the highest-scorers out of 425,000 adolescents taking the SAT (usually <13yo) starting in 1981, representing the top 0.01% of the test-takers. The TIP sample parallels the better-known SMPY sample, which also selected extremely intelligent adolescents, who were included in a longitudinal sample. It’s frequently suggested, based on anecdotal evidence or some biased convenience samples, that more intelligence may not be better; extremely intelligent people may be unhealthy, neurotic, insane, isolated, lonely, discriminated against by society and their peers, and doomed to failure; or if things are not quite that dire, as all studies show things improving up to 130, then at around that point greater intelligence may stop making any difference, and there be little difference between someone with an IQ of 130 and 160. This is difficult to study cross-sectionally, because once you start talking about as extreme as 0.01%, it is difficult to recruit any subjects at all, and your sample will be biased in unknown ways; if you only look at successful people, you are missing the hypothetical homeless bum living out of a trash can who is a troubled and misunderstood genius. To solve these problems, you want to filter through hundreds of thousands of people so you can select the very brightest possible, and you want to find them as early as possible in life, before they have had any chance to fail or succeed, and track them longitudinally as they grow up. This is what the SMPY & TIP studies do, and the results are that the subjects are spectacularly successful in life; great intelligence is not harmful and the returns to greater intelligence are not zero even as high as 1 in 10,000.

Makel et al 2016 also reports the ethnic breakdown of the TIP and SMPY samples: 72% white, 22% Asian, 6% not reported or other. This distribution might seem remarkable given that subjects taking the SAT in 1981 were born ~1970, when the USA was ~77% white, ~11% black, and ~0.7% Asian, so white are slightly under-represented, blacks are very under-represented (even if we assume all 6% are black, then that’s still half), and Asians are 31x (!) overrepresented.

## TIP/SMPY sample size & ethnic percentages: https://pbs.twimg.com/media/Cj9DXwxWEAEaQYk.jpg
tip <- 259; smpy <- 320 ## total: 579
white <- ((0.65*tip) + (0.78*smpy)) / (tip+smpy)
asian <- ((0.24*tip) + (0.20*smpy)) / (tip+smpy)
white; asian
# [1] 0.7218480138
# [1] 0.2178929188

# http://drjamesthompson.blogspot.com/2016/06/some-characteristics-of-eminent-persons.html
# > The data on ethnicity are rather sparse, but we can do a little bit of work on them by looking at US Census
# > figures for the 1970s when most of these children were born: White 178,119,221...Asia 1,526,401...So, in the
# > absence of more detailed particulars about the Other category, Asians win the race by a country mile. If we
# > simplify things by considering only Whites, Blacks and Asians the US in 1970 then the country at that time was
# > 88% White, 11% Black, and less than 1% Asian. The actual results of eminent students are 77% White, 0% Black,
# > 22% Asian. No need for a Chi square.
#
# Asian is 0.7%: R> 1526401  / (178119221 / 0.80)
whiteRR <- white / 0.77; asianRR <- asian / 0.007
whiteRR; asianRR
# [1] 0.937464953
# [1] 31.12755983

Of course, races in the USA have long differed by mean intelligence, with the rule of thumb being Asians ~105 IQ, whites ~100, and blacks ~90. So the order is expected - but still, 31x! Are the results being driven by some sort of pro-Asian bias or otherwise bizarre?

But this is an extreme sample. 1-in-10,000 is far out on the tails: 3.71SDs.

-qnorm(1/10000)
# [1] 3.719016485

Maybe this is normal. Can we work backwards from the overrepresentations to what differences would have generated them?

Yes, we can, even with this small sample which is so extreme and unrepresentative of the general population. This is because it is an order statistics problem: we know the order represented by the sample and so can work back to parameters of the distribution the order statistics are being generated by. Since IQ is a normal distribution, we know the overrepresentation RR, and the exact cutoff/limit used in the sample, we can convert the limit to a standard deviations, and then find the normal distribution $\mathcal{N}(100+x,15)$ which is RR (31) times the normal distribution $\mathcal{N}(100,15)$ at a standard deviations.

We can compare using two pnorms and shifting the second by a SDs. So for example, shifting by 15 IQ points or 1 SD would lead to 84x overrepresentation

pnorm(qnorm(1/10000)) / pnorm(qnorm(1/10000) - (15/15))
# [1] 84.39259519

We would like to solve for the shift which leads to an exact overrepresentation like 31.127; an optimization routine like R’s optim function can do that, but it requires an error to minimize, so minimizing pnorm()/pnorm(x) doesn’t work since it just leads to negative infinity, nor will RR == pnorm()/pnorm(x) work, because it evaluates to 0 for all values of x except the exact right one . Instead, we minimize the squared error between the ratio predicted by a particular x and our observed RR. This works:

## An optimization routine which automatically finds for us the IQ increase which most closely matches the RR:
solver <- function(RR, cutoff=10000) {
optim(1,
function(IQ_gain) { (RR - (pnorm(qnorm(1/cutoff)) / pnorm(qnorm(1/cutoff)-(IQ_gain/15))))^2 },
)$par } 100 + solver(whiteRR) # [1] 99.75488281 100 + solver(asianRR) # [1] 111.8929688 So our inferred white & Asian populations means are: 99.8 and 111.9. These are relatively close to the expected values. This approach can be used to infer other things as well. For example, the TIP/SMPY papers have not, as far as I’ve seen, mentioned what fraction of the white subjects were ethnic Jewish; since they are so over-represented in areas like Nobel prizes, we would expect many of the TIP/SMPY white students to have been Jewish. Using an estimate of the Jewish population in 1970 and estimates of their mean IQ, we can work forward to what fraction of TIP/SMPY subjects might be Jewish. The 1970-1971 National Jewish Population Study estimated 5,800,000 persons (of whom 5,370,000 were Jews) living in Jewish households out of a total US population of 205 million, or 2.8% of the total population or ~3.6% of the white population. So of the ~418 white subjects, ~15 would be expected to be Jewish under the null hypothesis of no difference. The majority of American Jews are of Ashkenazi descent5, for whom intelligence estimates are debated but tend to range 105-115 (with occasional samples suggesting even higher values, like Levinson 1957). In the Barbe 1964 Ohio sample (IQ ~143), 8% were Jewish6; in Terman’s (ratio IQ >140) 1920s sample in SF/LA, 10% were Jewish; Hollingworth’s 1930s sample (>180) turned up 51/55 or 90% Jewish7; Byrns 1936’s 1931 Wisconsin state sample found 18% of the Jewish sample to be in the top decile vs 10% American; in the Hunter College Elementary School sample 1948-1960 (>140, mean 157) in New York City, 62% were Jewish (Subotnik et al 1989, Subotnik et al 19938). Given estimates of the Jewish population of children in those specific times and places, one could work backwards to estimate a Jewish mean. We can calculate the fraction of the white sample being Jewish for each possible mean IQ: proportion <- function (gain, cutoff=10000) { (pnorm(qnorm(1/cutoff)) / pnorm(qnorm(1/cutoff)-(gain/15))) } possibleIQs <- seq(5, 15, by=0.5) data.frame(Advantage=possibleIQs, Fraction.of.white=(sapply(possibleIQs, proportion) * 15) / 418) Advantage Fraction.of.white 1 5.0 0.1415427303 2 5.5 0.1633099334 3 6.0 0.1886246225 4 6.5 0.2180947374 5 7.0 0.2524371552 6 7.5 0.2924980125 7 8.0 0.3392769622 8 8.5 0.3939561508 9 9.0 0.4579348680 10 9.5 0.5328710150 11 10.0 0.6207307813 12 10.5 0.7238482059 13 11.0 0.8449966589 14 11.5 0.9874747049 15 12.0 1.1552093388 16 12.5 1.3528802227 17 13.0 1.5860693342 18 13.5 1.8614413902 19 14.0 2.1869615788 20 14.5 2.5721585555 21 15.0 3.0284424112 Judging from earlier samples with very high cutoffs, I’d guess TIP/SMPY has at least a majority Jewish, giving a mean IQ of ~110; this is pleasantly similar to estimates based on regular samples & estimation. This result is also similar to La Griffe du Lion’s 2003 threshold analysis estimating a mean IQ of 112 based on Ashkenazi overrepresentation among USSR championship chess players, 111 based on Western Fields Medal awards, and 110 based on the USA/Canada Putnam competition. But if the mean IQ was as high as 112, then almost every single white subject would be Jewish in every sampling, which seems implausible and like something so striking that anyone writing or involved with TIP/SMPY would have to have mentioned at some point - right? For the same reason, the original estimate of 112 for the Asians strikes me as on the high side. This could be due to problems in the data like underestimating the Asian population at the time - perhaps the Southeast/Midwest states that TIP samples from were more than 0.7% Asian - or it could be due to sampling error (only n=579, after all). Working backwards doesn’t immediately provide any measurement of precision or confidence intervals. Presumably someone has worked out analytic formulas which come with standard errors and confidence intervals, but I don’t know it. Instead, since the selection process which generated our data is straightforward (population mean -> millions of samples -> take top 1-in-10000s -> calculate overrepresentation), I can again use Approximate Bayesian computation (ABC) to turn a simulation of the data generating process into a method of Bayesian inference on the unknown parameters (population means) and get credible intervals. What sort of confidence do we have in these estimates given that these RRs are based only on? We can simulate TIP/SMPY-like selection by taking the hypothetical means of the two groups, generating ~3 million simulates (579 * 10000) each, selecting the top 1/10000th, taking the RRs and then solving for the mean IQ. If we provide a prior on the means and we hold onto only the means which successfully generate TIP/SMPY-like fractions of 72% & 21%, this becomes ABC with the saved means forming the posterior distribution of means. (It would likely be faster to use MCMC like JAGS, but while JAGS provides truncated normal distributions which one could sample from quickly, and the necessary pnorm/qnorm functions, but it’s not clear to me how one could go about estimating the overperformance ratio and the binomial.9) For my priors, I believe that the rule of thumbs of 100/105 are accurate and highly unlikely to be more than a few points off, so I use a very weak prior of populations means being $\mathcal{N}(100/105, 4)$. In exact ABC, we would keep only data which exactly matched 72%/22%, but that would require rejecting an extremely large number of samples. Here we’ll loosen it to ±2% tolerance: simulateTIPSMPY <- function() { ## informative priors: IQs are somewhere close to where we would estimate based on other datasets whiteMean <- round(rnorm(1, mean=100, sd=5), digits=2) asianMean <- round(rnorm(1, mean=105, sd=5), digits=2) iqCutoff <- 100 + -qnorm(1/10000) * 15 whites <- rnorm(0.770 * 579 * 10000, mean=whiteMean, sd=15) whiteSample <- max(1, sum(ifelse(whites>iqCutoff, 1, 0))) asians <- rnorm(0.007 * 579 * 10000, mean=asianMean, sd=15) asianSample <- max(1, sum(ifelse(asians>iqCutoff, 1, 0))) ## white+Asian = 92% of original total sample, so inflate by that much to preserve proportions: 1.08 totalSample <- (whiteSample+asianSample) * (1 + (1-(white+asian))) whiteFraction <- round(whiteSample / totalSample, digits=2) asianFraction <- round(asianSample / totalSample, digits=2) # print(paste("samples: ", c(whiteSample, asianSample), "fractions: ", c(whiteFraction, asianFraction))) tolerance <- 0.02 if ((abs(whiteFraction - 0.7218480138) < tolerance) && (abs(asianFraction - 0.2178929188) < tolerance)) { return(data.frame(White=whiteMean, Asian=asianMean)) } } library(parallel); library(plyr) simulateSamples <- function(n.sample=10000, iters=getOption("mc.cores")) { ## because of rejection sampling, no run is guaranteed to produce a sample so we loop: results <- data.frame() while (nrow(results) < n.sample) { simResults <- ldply(mclapply(1:iters, function(i) { simulateTIPSMPY() } )) results <- rbind(results, simResults) # print(paste("Samples so far: ", nrow(results))) } return(results) } posteriorSamples <- simulateSamples() mean(posteriorSamples$White < posteriorSamples$Asian) # [1] 1 ## we have relatively few samples, so get a better posterior estimate by shuffling the posterior samples & comparing many times: mean(replicate(1000, mean(c(sample(posteriorSamples$White) < sample(posteriorSamples$Asian))))) # [1] 0.9968822 quantile(probs=c(0.025, 0.975), posteriorSamples$White, na.rm=TRUE)
#     2.5%     97.5%
# 89.49975 101.38050
quantile(probs=c(0.025, 0.975), posteriorSamples$Asian, na.rm=TRUE) # 2.5% 97.5% # 101.37000 116.74075 par(mfrow=c(2,1)) hist(posteriorSamples$White, main="Posterior white mean IQ estimated from TIP/SMPY cutoff & ratio", xlab="IQ")
hist(posteriorSamples$Asian, main="Posterior Asian mean", xlab="IQ") So sampling error does turn out to be substantial: our 95% credible intervals are white 90-101, Asian 101-116. Still, the overlap is minimal, with P=99.7% that the Asian mean is higher than the white. We are able to conclude that the rank ordering is highly likely to be correct, and the results are consistent with the conventional wisdom, so there is no prima facie case for bias in the results: the ethnic composition is in line with what one would calculate from the design of TIP/SMPY and population means. Genius Revisited: On the Value of High IQ Elementary Schools Genius Revisited documents the longitudinal results of a high-IQ/gifted-and-talented elementary school; one of the most striking results is the general high education & income levels, but absence of great accomplishment on a national or global scale (eg a Nobel). The authors suggest that this may reflect harmful educational practices at the elementary school or the low predictive value of IQ. I suggest that their standards fall prey to a base-rate fallacy, and that the lack of accomplishment is inherent and unavoidable as it is driven by the regression to the mean caused by the relatively low correlation of early childhood & adult IQs; in contrast, its associated high-IQ/gifted-and-talented high school, which has access to more predictive IQ scores, has much higher achievement in proportion to its lesser regression to the mean (despite dilution by elementary students being automatically enrolled). This greater-regression problem can be lessened by the use of additional variables in admissions, such as parental IQs or high-quality genetic polygenic scores; unfortunately, these are either politically unacceptable or dependent on future scientific advances. This suggests that such elementary schools may not be a good use of resources or high school slots. Genius Revisited: High IQ Children Grown Up, by Subotnik, Kassan, Summers & Wasser 1993 is a short (142 pages) book reporting the results of a longitudinal/followup study in 1988 of 210 of the 600 1948-1960 alumni of the Hunter College Elementary School (HCES) who had reached their 40s or so. (See also the brief more statistically-oriented report of the survey results in High IQ children at midlife: An investigation into the generalizability of Terman’s genetic studies of genius, Subotnik et al 1989; for a overview of gifted education with some mention of the HCES results, see Subotnik et al 2011.) Hunter Elementary is a small elementary school in New York City enrolling ~50 students each year starting in preschool/kindergarten since the 1940s, who then typically enroll in the associated Hunter College High School, itself associated with Hunter College. Hunter Elementary is famous for extremely stringent admission based on IQ tests, yielding a student body with a mean IQ in the 150s (or around 1-in-10,000); the gifted students are taught a wide-ranging and enriched curriculum designed for gifted children. (If you’ve ever read about helicopter or tiger moms in Manhattan training their kids on IQ tests to get them into an elite kindergarten, Hunter Elementary is one of the kindergartens they have in mind.) As such, Hunter Elementary students might be expected to be extremely interesting and highlight the effects of great intelligence on one’s life: as they all are selected young and relatively systematically from NYC children, such a longitudinal study is going to be much more reliable than other attempts at studying high intelligence using cross-sectional or ad hoc recruitment from child psychologists. High IQ background Parallel to the Hunter Elementary students, but much better known are the Terman study (young, relatively low IQ), Anne Roe’s studies of world-class scientists (generally in their 40s or 50s), and the SMPY and TIP longitudinal studies (almost identical cutoffs but measured in middle school ~12yo using SATs, similar to Hunter High admission); some relevant publications: I came across Genius Revisited while looking into the question of inferring ethnic composition of the SMPY/TIP samples based on the high cutoff threshold and becoming intrigued by a mention by Charles Murray in his article Jewish Genius of a NYC elementary school with mean IQ >150 where 24 of the 28 highest scoring students were Jewish, an elementary school I didn’t remember ever seeing mentioned in discussions of high IQs/life outcomes, and ordered a copy. Aside from trying to track down a reference for Murray’s Jewish claim (which turned out to not be mentioned in the book aside from the overall Jewish percentage), while a high school for the gifted makes sense, I had some doubts about whether such an elementary school made sense and was curious how it had turned out. HCES results To summarize the results: contrary to stereotypes that bookish, nerdy, socially inept, absentminded, emotionally dense, arrogant and unfriendly, and that they are loners, high IQ children are physically and psychologically healthy, if not healthier; they are often socially capable; adult accomplishment and eminence increase with greater intelligence, with no particular cutoff visible at places like IQ 130; happiness is not particularly greater; male and female differences in achievement exist but are at least partially driven by other sex-linked differences in choice of field and work-life balance; particular ethnicities are under or overrepresented as one would calculate using the normal distribution from the study-specific cutoffs and ethnic means; and even with extremely high ability in all areas, people tend to eventually specialize in their greatest strength which is their comparative advantage; and overall educational credentials are much more common in the later groups than the earlier ones. So what does Genius Revisited report? In general, it is surprisingly light on detailed quantification or analysis. Income and education are reported only cursorily; adult achievements are not gone into any sort of detail or categorization only than vague generalizations about there being lots of doctors, professors, and executives etc. They do not report adult IQs, or attempt any statistical analysis to compare IQs at admittance, graduation, or when contacted as adults, whether some subtests predict adult accomplishment better than others, whether there were differential regressions to the mean or whether there was any regression to the mean observed by graduation10, or comparison of any dropouts/transfers with the students who graduated Hunter Elementary and continued to Hunter High; the questionnaires are based on the old Terman questionnaires and don’t seem well focused to investigate modern concerns in gifted education or individual differences psychology. From this perspective, the book is quite a disappointment, as there are not many high IQ longitudinal datasets around - yet they waste the opportunity. Some further details and more fine-grained categorization of a few of the hundred variables collected are reported in Subotnik et al 1989 but the treatment is much less than it could have been. What it does do is attempt as a sort of narrative ethnography by piecing together many quotes from the students about their Hunter Elementary experience and later life. This is interesting to me on a personal level because my parents had considered sending me to the Long Island School for the Gifted but ultimately decided against it; so in a way, reading their memories is a glimpse of a path not taken. The picture that emerges confirms in many respects the portrait of children in Terman/SMPY/TIP: the children are healthy, well-socialized, enjoy outdoor sports (particularly hiking); girls tend to not prefer the stereotypical childhood activities like dolls (which is interesting given SMPY results related to testosterone); reading is, of course, everyone’s favorite hobby, especially to help with researching their other hobbies; the burden of being labeled a genius or prodigy bothered some but apparently not most of them; students remembered Hunter Elementary extremely fondly and were glad to have gone there rather than regular school, although opinions on how Hunter Elementary could have been better are amusingly equally divided in Subotnik et al’s recounting (a good compromise leaves everyone unhappy); teachers likewise regarded teaching there as a plum assignment, as the students were highly cooperative, enthusiastic, almost always well-behaved, soaked up material like sponges, and would happily go off on tangents like debating the strategic value of Australia during WWII (in other words, what any would-be teacher dreams of teaching, instead of getting a class of bored, sleepy kids who act out and forget things the second you explain them); many students deliberately did not pursue the most demanding adult careers to have a work-life balance, particularly the women, with the usual differences in subject-area preferences; women were, as predicted given the later era than Terman, far more likely to pursue higher education and some sort of employment; students are highly successful, but none seemed particularly extraordinarily successful. There is also a short comparison with Hunter Elementary in the 1990s; apparently much the same as in the 1960s, with the main interesting change that Hunter Elementary has added a racial quota for black students, but Subotnik et al claim that the mean IQ scores have not fallen substantially. It would be interesting to know exactly how much it has fallen, how many of the black students have immigrant parents, and how many students are now of East Asian descent.11 Overall, the writing is clear and there is, if anything, insufficient technical jargon. Some dry humor appears in spots (eg in Subotnik et al 1989, a wry comment on rent control and the difficulties of longitudinal studies: the only addresses on file were those of the parents while the child attended the school. Fortunately, given the state of the New York City housing market, checking those addresses against the 1988 Manhattan phone book proved to be fairly productive). Disappointingly average Subotnik et al generally seem to hold what has been called the resource model of gifted education: high IQ children have much better odds of growing up into the great movers & shakers and thinkers of the world who have disproportionate influence on what happens (definitely); that special measures such as enriched education, schools with peers in intelligence, and accelerated courses will increase the yield of great (maybe); and that the increase justifies the upfront expenses (uncertain). By success, they have very high standards; Gallagher’s foreword speaks for the rest of the book when it says: The authors were disappointed to discover that although this sample succeeded admirably in traditional terms, with its share of physicians, lawyers, and professors, there were no creative rebels to shake society out of its complacency or revolutionize a field. Further: Norbert Wiener, in his book The Autobiography of an Ex-Genius [Ex-Prodigy: My Childhood and Youth], detailed his unhappy family life with a domineering father and enough personal problems to be in and out of mental institutions. Yet, it was this Norbert Wiener who gave the world cybernetics that revolutionized our society. What if he had had a happy family life with a warm and agreeable father? One is left to wonder whether Wiener would have had the drive and motivation to make this unique contribution. The same question can be posed for these Hunter College Elementary School graduates. Are many of them too satisfied, too willing to accept the superior rewards that their ability and opportunity have provided for them? What more could they have accomplished if they had a psychological worm eating inside them - whether that worm was low self-concept or a need to prove something to someone or to the world - that would have driven these people to greater efforts. What if their aptitudes had been challenged in a more hard-driving manner, like Wiener’s experience, into the development of a specific talent? This book raises many significant, sometimes disturbing issues… The authors raise some disturbing issues regarding the purposes of schools for the gifted. Indeed, just what is the contemporary rationale for funding schools or programs for the highly gifted student? If one is looking to such an institution as a source of leading students towards societal leadership (or, as the authors suggest, a path to eminence), then the Hunter College Elementary School of the past failed to realize such an aspiration. Indeed, this goal may well be beyond the reach of any elementary school…the [Hunter College] High School seeks to enhance students’ commitment to intellectual rigor and growth, develop opportunities for specialization, and commitment to caring and compassion. Will such a rationale foster more students down the path towards genius? The research literature and the current study would indicate that such a condition is a necessary but not sufficient condition to move students into making ground-breaking discoveries or toward professional eminence. Does it follow then that such schools should not exist? Or at least, not at public expense? I would vigorously argue against both reactions. This fits with the general description of the Hunter Elementary cohort on pg3-4: The mean IQ of the Hunter sample was 157, or approximately 3.5 standard deviations above the mean, with a range of 122 to 196 on the L-M form. …Each class at Hunter College Elementary School from the years 1948 to 1960 contained about 50 students, yielding a total possible population of 600 graduates…35% of the total population of 1948-1960 HCES students (n=210) completed and returned study questionnaires Religious Affiliation: The Hunter group is approximately 62% Jewish, although they describe themselves as Jews more in terms of ethnic identity than religious practice. The group, as a whole, is not religious. Educational Attainments: Over 80% of the study participants held at least a Master’s degree. Furthermore, 40% of the women and 68% of the men held either a Ph.D, LL.B., J.D., or M.D. degree. Occupation and Income: Only two of the HCES women identified themselves primarily as homemakers. 53% were professionals, working as a teacher at the college or pre-college level, writer (journalist, author, editor), or psychologist. The same proportion of HCES men were professionals, serving as lawyers, medical doctors, or college teachers. The median income for men in 1988 was$75,000 (range = $500,000) and for women$40,000 (range = $169,000). Income levels were significantly different for men and women, even when matched by profession. For example, the median income for male college teachers or psychologists was$50,000 and for females, $30,000 By regular standards, this is a remarkably high degree of accomplishment. Even now, only a small fraction of the population can be said to hold a Ph.D, LL.B., J.D., or M.D., but in the Hunter Elementary cohort, you could hardly throw a rock without hitting a professor (16% of men), who would then be able to turn to the person standing next to them to have their wound treated (18% doctors), and turn to the person on the other side in order to sue you for assault (20% lawyers). For this cohort, the education baseline would be more like <7%, not >80%. Subotnik et al 1989 breaks it down a little more precisely in Table 2 Highest Degree Attained: for men, 4% not available, 20% Bachelors, 43% Masters, 40% Ph.D/L.L.B./J.D./M.D. The income levels are also sky-high: in 1988, median household income would’ve been ~$50,000, and the ranges like $500,000 indicate that Hunter Elementary incomes stem from life choices and career preferences as much as any limits from ability. But it doesn’t fit the definition of great accomplishments. They mention no one winning a Nobel, or a Pulitzer, or being globally famous. Thus, in a real sense, Hunter Elementary has failed, and with it (the authors imply), the idea that IQ is the driving force behind greatness; thus, Subotnik et al spend much of the book, and other publications, pondering what is missing. If IQ is merely a necessary factor or threshold, but one that still leaves such a high chance of an ordinary life, what really makes the difference? Is the crucial ingredient a drive for mastery? Did Hunter Elementary accidentally quash students’ ambitions for a lifetime by de-emphasizing competition and grades? Or (as the other half of surveyed students maintained), did it have too much competition and broke the students mentally? Was Hunter Elementary too well-equipped a cocoon, leaving students unprepared for Hunter High and the real world, or not enough? Did the home environment determine this, or the curriculum? Did the broad academic curriculum leave students a mile wide and an inch deep and lacking in fundamentals acquired by drilling and repetition? Sample size But should we declare it a failure, considering the parallel lines of evidence from Roe, SMPY, and TIP? The mentioned standard is a high bar indeed. What percentage of the population can be truly said to revolutionize a field? It’s a lifetime’s work just to truly understand a field and reach the research frontier and make a meaningful contribution, and most of the population generally doesn’t even try but pursue other goals. Out of 600 students, is it reasonable to consider the Hunter Elementary experiment a failure because none has (yet - the Nobel Prize is increasingly delayed by decades)? As Gallagher then points out: …Yet, there are very few such individuals alive in any particular era. The statistical odds against any one of them having graduated from one elementary school in New York City is great. Whether the creative rebel would have survived the selection process at Hunter, or any similar school, is one of those remaining questions that should puzzle and intrigue us. If we consider the STEM Nobel Prizes, the USA has perhaps 1 per million people. So if even 1 HCES student had won a STEM Nobel out of 210, or 600, that would imply an enormous increase in odds ratio of >1666 ($\frac{\frac{1}{1}}{\frac{600}{1000000}}$); or to put it another, if we genuinely expected 1 or more Nobels from our HCES alumni, then to achieve that >1666 increases in odds with only +57 early-childhood IQ points, we’d also have to believe something along the lines of each individual IQ point on average increasing the odds by 29x! And of course, if we did believe in such effect sizes, we would still frequently expect to observe a HCES-sized cohort to not win a Nobel (eg if we had expected 1 Nobel prize per 600, for a probability of $\frac{1}{600}$ per student, then the probability of seeing 0 Nobels in n=600 is very high: $(1-\frac{1}{600})^{600} = 0.367$; to drive the non-Nobel probability down to <5%, we would have to expect >=3 Nobels per 600). So it’s unclear how much weight we ought to put on the apparent failure of the HCES alumni, because even the ludicrously optimistic model is consistent with often seeing failure. Weak childhood IQ scores: regression to the mean How many people from Hunter Elementary come anywhere close? If we were to double-check in Wikipedia by looking for Notable people whose entries link to Hunter Elementary, perhaps because they were students there, we find painter Margaret Lefranc, linguist E. Adelaide Hahn, and minor actor Fred Melamed, and Supreme Court justice Elena Kagan (but while her mother taught at Hunter Elementary, she herself went to Hunter College High School - along with at least 95 other Notable people). I later learned that Hamilton star Lin-Manuel Miranda and scientist Adam Cohen also went to Hunter Elementary as well as High. Triple-checking in Google, this does seem to be a fair accounting - no billionaires or Nobelists suddenly pop out. If we were to judge by Wikipedia entries, it would seem that Hunter Elementary can claim around 5 Notable alumni while Hunter High can claim 96. (Checking the 96 WP entries by hand, most omit mention of the elementary school or whether they passed exams to get into Hunter High, but the ones who do always specify exams or a non-Hunter Elementary; only 1 entry, the group entry for the hip-hop band Dujeous, turns out to include a Hunter Elementary member: Loren Hammonds/Mojo the Cinematic. Overall, this comparison may be somewhat biased against Hunter Elementary but I don’t think hugely so.) This is not because Hunter High is 32x larger than Hunter Elementary: Hunter Elementary currently accepts ~50 students per year while Hunter High currently accepts ~175 + 50 grandfathered in from Hunter Elementary (total ~225), and is only 4.5x bigger - 3.5x if we exclude the Hunter Elementary alums (who do not appear in the 95+ listed, apparently). Even more strikingly, while I do not recognize the names of Lefranc, Hahn, Melamed, or Adam Cohen, I do recognize several names on the Hunter High list (Kagan, of course, but also Bruce Schneier, Mark Jason Dominus, some rappers in passing). This would imply that Hunter High grads are much more likely to achieve Notability than Hunter Elementary grads: something like 8 times more likely. Why? Another way would be to ask what should we expect, from a statistical and psychometric point of view, from Hunter Elementary students, given the procedures and tests used? There are a number of statistical issues which can arise in intelligence research particularly: range restriction such as ceiling/floor effects, measurement error biasing correlations down and requiring correction, sampling error, loss of measurement invariance in IQ tests or test-specific learning leading to hollow gains (particularly prevalent in interventions), genetic confounding of correlations between IQ and other variables like SES, test-retest reliability, and so on. (Many of these are discussed in more detail in Hunter & Schmidt’s 2004 textbook Methods of Meta-analysis: Correcting Error & Bias in Research Findings.) As Hunter Elementary used and still uses a legitimate IQ test (Stanford-Binet Intelligence Scales), the results are not interventional or claimed to be causal, and we are concerned with them as a group compared to the general population, the last issue of reliability/predictive validity is the one which bothers me the most in trying to interpret the results. Hunter Elementary uses IQ testing of ~5yo children, selecting those >IQ 140 and getting a mean of IQ 157 (3.8 SDs); these children are then kept enrolled in Hunter Elementary and grandfathered into Hunter High as long as their grades stay reasonable, with expulsions and transfers apparently rare (and little mentioned in the book). However, as is well known, childhood IQs are imperfect predictors of final adult IQs, for various neurological, developmental, and genetic reasons; the best possible estimate at 5yo will still only correlate with adult IQ at perhaps r=0.5-0.6. And having been selected for scoring extremely high on a particular test, Hunter Elementary kids must revert to mediocrity. What can we estimate their adult IQs to be? Since the majority of students are Jewish (or these days, split between those of Jewish and East Asian descent) whose mean is usually estimated at something like 110, we could predict that their adult IQs will not average 157, but will average 110 + (157-110) * 0.5 = 133. 133 IQ is nothing to sneeze at, but it is also only +2.2SDs and closer to 1 in 50 than 1-in-10,000; a Hunter Elementary school grad could easily not even qualify for MENSA. Or to put it another way, with 260 million people in the USA in 1993, there were around 3.6 million people with IQs >=133, of which the total Hunter Elementary cohort would represent 0.016%. If we consider cohorts of 600 children with adult mean IQs of 133, not many of them will be >157 at all - only 5% or ~32 students (mean(replicate(100000, sum(sort(rnorm(600, mean=133, sd=15))>157)))). The others will have developed into adult IQs below that, possibly much below that. This calculation doesn’t require any knowledge of outcomes and could have been done before Hunter Elementary opened: inherently, due to the limits of IQ tests in screening for extremely gifted adults based on noisy early childhood tests, most positives will be false positives. (This is the same as the famous mammography and terrorist screening examples of how an accurate test + low base-rate = surprisingly high false positive rate and low posterior probability.) What about Hunter High? Hunter High tests sixth graders who enroll as 7th graders; 6th graders tend to be ~11yo, not 4-5yo. One correlation quoted by Eysenck is testing 11yos can have a correlation of ~0.95 with adult scores; so Hunter High grads, assuming they had the same mean (I haven’t seen any means quoted), would expect to revert to mediocrity down to 110 + (157-110) * 0.95 = 154 ie almost identical. (With a correlation of 0.9, 152, and so on). So out of 600 Hunter High alums, 252 will remain >157, or ~8x the Hunter Elementary rate. That is, the overrepresentation of Hunter High graduates among Hunter-related Notable figures is almost identical to their overrepresentation among Hunter-related graduates who maintain their elite IQ status. None of the materials I have read on Hunter Elementary, aside from one article in New York magazine12 drawing on Lohman & Korb 2006’s Gifted Today but Not Tomorrow? Longitudinal changes in ability and achievement during elementary school, have mentioned the issue that IQ tests in such early childhood are simply not that predictive in finding extreme tails, or even alluded to it as a problem, so I have to wonder if Subotnik et al13 appreciate this point: from basic psychometric principles, we would predict that Hunter Elementary graduates will not be extraordinarily intelligent, will represent only the tiniest fraction of the population of intelligent people, and thus their adult accomplishment will not be out of line with what we observe - solid academic and social achievement. Nor is there any particular reason to attribute their failure to the atmosphere or curriculum or methods of Hunter Elementary itself. Implications for gifted education Given this, we would have to conclude that the idea of a gifted & talented elementary school is difficult to justify on any grounds related to focusing resources on students’ with future adult intelligence >150 as only a few percent of such students are findable with current IQ testing methods at that age, but that it makes far more sense to screen at a later age like 11yo and concentrate resources at high school or college levels. If we concluded that the gain from better education of those 5% in an elementary school is profitable and so a Hunter-like elementary school is a good idea, we should definitely not automatically enroll all such elementary school students in a even more expensive Hunter-like high school: each such grandfathered student is worth ~1/8th an outsider student in terms of potential. It would be much better to not grandfather the elementary school students - they have already been highly advantaged by the enriched education & peers, after all, so why should they be given an additional huge advantage over all the students outside the system who are equally deserving of the chance? The main reason would seem to be some sort of family or loyalty sentimental reasoning; if this bias cannot be overcome, the idea of a single vertically integrated feeder system may be actively harmful to gifted education. Matters could be improved, though, with more broad-ranging tests. For example, genetics: as adult IQ is a highly heritable trait with perhaps up to 80% of variance predictable from all genetic variants and >~50% predictable from all SNPs, with the heritability increasing with age and only ~25% at age 5 (Bouchard 2013), predictions of adult IQ based on 5yo testing could be improved substantially using their parents’ & siblings’ IQs, or by direct genetic prediction; this would help identify the children who are rejected because of developmental quirks but who would eventually live up to their genetic potential. If we consider a path model with genes ~> IQ (0.50), IQ5yo ~>IQ (0.50), genes ~> IQ5yo (0.25): model <- 'IQ_adult ~ 0.8*Gene + 0.5*IQ_5 IQ_5 ~ 0.25*Gene' d <- simulateData(model) s <- sem(model, std.ov=TRUE, data=d) semPaths(s, "Standardized", "Estimates", style="lisrel", curve=0.8, nCharNodes=0, edge.color="black", label.scale=FALSE, residuals=FALSE, fixedStyle=1, freeStyle=1, exoVar=FALSE, sizeMan=10, sizeLat=24, label.cex=3, edge.label.cex = 2.2) Then using an ideal SNP genetic score and a 5yo IQ test, one could expect to predict 0.5 + (1-0.25)*0.5 = 0.875 or 87% of variance, giving a prediction/adult IQ of sqrt(0.875) = 0.93; with this sort of predictive power, the reversion to mediocrity is minimal and Hunter Elementary kids would then have adult IQs of 110 + (157-110) * 0.93 = 153. In that scenario, we could create a Hunter-like Elementary school which is as good at filtering as Hunter High is. While it’s unclear when we will be able to predict 50% of variance in adult IQs based on polygenic scores, in the near future we can hope for polygenic scores on the order of 10%, which would still be helpful: polygenicscore = 0.10; 110 + (157-110) * sqrt(polygenicscore + (1-0.25)*0.5) = 142.4. Besides waiting for better polygenic scores, other factors could be included in a predictive model such as parental IQs and income/education, sibling IQs, and race. I don’t know if such an elementary school for the gifted would be feasible, however: improved predictions will still maintain or increase the existing controversial racial disparities, the selection may strike the public as even more unfair than it is now, and will inherently yield classrooms with more cognitive inequality at the moment (despite how the slower students will tend to catch up with time) which may itself impede the educational mission or foster resentment & rivalry. Ultimately, it would seem that the most justifiable reason for running Hunter Elementary is the reason that comes across most clearly reading the alumni reminiscences: because they would have been miserable in regular schools. If early-developing children must be subjected to mandatory education, then it should at least be with their peers. Great Scott! Personal Name Collisions and the Birthday Paradox How large does can a social circle be before first names no longer suffice for identification? Scott, I’m looking at you.MakerOfDecisions, 29 July 2016 Scott here refers to any of Scott Alexander, Scott Adams, Scott Aaronson, Scott Sumner (and to a much lesser extent, Scott Garrabrant, Orson Scott Card, and Scott H. Young); a reference to a Scott on a site like Less Wrong is increasingly ambiguous. When a large number of samples draw from a common pool of identifiers, collisions are common, leading to the birthday paradox: despite there being 365.25 days in the year, a classroom of just 23 people (who can cover at most 6% of the days in a year) is ~50% likely to have at least two people who share the same birthday and so birthdays cease being unique unambiguous identifiers. (Intuitively, you might expect the number to be much larger and closer to 180 than 23.) We can verify this by simulation: dupes <- function(a) { length(a) != length(unique(a)) } identifiers <- function(n, ids, probabilities) { sample(1:ids, n, prob=probabilities, replace=TRUE) } simulate <- function(n, ids, probabilities=rep(1/ids, ids), iters=10000) { sims <- replicate(iters, { id <- identifiers(n, ids, probabilities) return(dupes(id)) }) return(mean(sims)) } simulate(23, 365) # [1] 0.488 sapply(1:50, function(n) { simulate(n, 365) } ) # [1] 0.0000 0.0029 0.0059 0.0148 0.0253 0.0400 0.0585 0.0753 0.0909 0.1196 0.1431 0.1689 0.1891 # 0.2310 0.2560 0.2779 0.3142 0.3500 0.3787 0.4206 0.4383 0.4681 0.5165 0.5455 0.5722 0.5935 # [27] 0.6227 0.6491 0.6766 0.7107 0.7305 0.7536 0.7818 0.7934 0.8206 0.8302 0.8465 0.8603 0.8746 # 0.8919 0.9040 0.9134 0.9248 0.9356 0.9408 0.9490 0.9535 0.9595 0.9623 0.9732 Similarly, in a group of people, it will be common for first names to overlap. How common? There are far more than 365.25 first names, especially as some first names are made up by parents. Names have a highly skewed (often said to be a power law) distribution: the first few baby names make up an enormous fraction of all names, hence all the Ethan/Lucas/Mason baby boys in 2016. (One would think that parents would go out of their way to avoid too-popular names, but apparently not.) Since there are only 10,000 things under heaven, one might think that the top 10000 personal names would give a good guess. At what n can we expect a collision? findN <- function(ids, targetP=0.5, startingN=1, probabilities=rep(1/ids, ids)) { n <- startingN collisionProbability <- 0 while (collisionProbability < targetP) { collisionProbability <- simulate(n, ids, probabilities) n <- n+1 } return(n) } findN(10000) # [1] 118 simulate(118, 10000) # [1] 0.5031 We could also use an approximation such as the square approximation: $n \approx \sqrt { 2m \times p(n)}$: sqrt(2 * 10000 * 0.5) ~> 100 Or the similar upper bound: ceiling(sqrt(2*10000*log(2))) ~> 118. So the collision point is smaller than Dunbar’s number. But all of these are themselves upper bounds because the case in which birthdays/names are uniformly distributed is the worst case. If there is any difference in the probabilities, a collision will happen much earlier. This makes sense since if 1 birthday happens with, say, P=0.99, then it’s almost impossible to go more than 3 or 4 birthdays without a collision. Likewise, if one birthday has P=0.50, and so on down to P=$:

sapply(1:23, function(n){ simulate(n, 365, probabilities=c(0.99, rep(0.01/364, 364)))})
# [1] 0.0000 0.9789 0.9995 1.0000 1.0000 1.0000 1.0000 1.0000 1.0000 1.0000 1.0000 1.0000 1.0000
#     1.0000 1.0000 1.0000 1.0000 1.0000 1.0000 1.0000 1.0000 1.0000 1.0000
sapply(1:23, function(n){ simulate(n, 365, probabilities=c(0.5, rep(0.5/364, 364)))})
# [1] 0.0000 0.2531 0.5031 0.6915 0.8182 0.8896 0.9402 0.9666 0.9808 0.9914 0.9951 0.9973 0.9988
#     0.9993 0.9991 0.9999 1.0000 1.0000 0.9999 1.0000 1.0000 1.0000 1.0000

How skewed are real names? Given Names Frequency Project provides Popular Given Names US, 1801-1999 (1990-1999, 909288 names) based on Social Security data. After deleting the first 4 lines of s1990m.txt, it can be loaded into R and the fractions used as probabilities to find the 50% collision point for US names:

names <- read.csv("s1990m.txt", header=FALSE)
summary(names)
#         V1            V2
#  Aaron   :  1   Min.   :   55.0000
#  Abdiel  :  1   1st Qu.:   86.0000
#  Abdullah:  1   Median :  183.0000
#  Abel    :  1   Mean   :  914.1923
#  Abraham :  1   3rd Qu.:  535.5000
#  Adam    :  1   Max.   :24435.0000
#  (Other) :852
sum(names$V2) # [1] 784377 ## "Scott" as fraction of all names: 2279 / 784377 # [1] 0.0029054906 ## presumably male names: 2279 / (784377*0.5) # [1] 0.005810981199 simulate(118, nrow(names), probabilities=names$V2/sum(names$V2)) # [1] 1 findN(nrow(names), probabilities=names$V2/sum(names$V2)) # [1] 15 So a more realistic analysis suggests n=15 is where unique first names will probably break down. This only covers the 853 most common personal names, and the more names, the higher the n has to be to trigger a collision (making 15 something of a lower upper bound); to estimate 10000, we need to fit a distribution to extrapolate below that. The log normal distribution fits reasonably well and is easy to work with: library(fitdistrplus) fitdist(names$V2, "lnorm")
# Fitting of the distribution ' lnorm ' by maximum likelihood
# Parameters:
#            estimate    Std. Error
# meanlog 5.550448321 0.04640182299
# sdlog   1.359185357 0.03281096378

simulateLN <- replicate(100, {
names <- rlnorm(10000, meanlog=5.550448321, sdlog=1.359185357)
hit <- findN(length(names), startingN=46, probabilities=names/sum(names))
return(hit)
})
median(simulateLN)
# [1] 51

Since names will cluster by age group, location, profession, and whatnot, arguably even 51 is a bit of an upper bound.

Finally, one might ask the probability of a group with a great Scott, or to put it another way, the probability of it unfortunately getting away scot-free.

This is easy to answer; the probability of having 1 or more Scotts in a group is the probability of everyone having a name other than Scott. We saw that the probability of being named Scott was 0.0029054906 in the name dataset. So the probability of one person not being named Scott is 1-0.0029054906 or 0.997. So the probability of n people all being named not-Scott is 0.997n. The crossover point is ~239.

So an American social group cannot exceed n=51 before first names begin to break down, and it is all Scott’s fault at n=239.

Detecting fake (human) Markov chain bots

Some popular Twitter and Tumblr accounts use Markov chains trained on a corpus of writing such as Markov James Mitchens or two unrelated corpuses to create amusing mashups: programming documentation and H.P. Lovecraft’s horror/SF fiction or the King James Bible or the works of Karl Marx, Kim Kardashian and Kierkegaard, or Silicon Valley recruiting emails and Erowid drug use reports. The humor comes from the fact that the Markov chains have no understanding and are merely programs producing gibberish that occasionally present striking juxtapositions or insights. Much of their appeal derives in large part from the fact that while humans curate them, humans don’t write them. They depend on this authenticity to be striking.

Of course, there’s always the temptation to edit them or write them wholesale, perhaps because the Markov chains aren’t cooperating in producing any comedy gold to tweet that day, which deceives the reader. This poses an inverse Turing test: how would you detect a fake Markov chain account, that is, one where a human is pretending to be a computer and writing some of the text?

Markov chains are trained on a specific corpus and are a probabilistic generative model which encode the probability that a word X follows another word Y for all the words in that corpus (and similarly if they are operating on letters or on n-grams); there is no state or memory or look back or ability to model recursion. To generate text, one simply picks a random word Y, looks up the probabilities of all the words AZ from Y, and picks a word at random weighted by those probabilities; then repeat indefinitely. Conversely, one could also use it to calculate the likelihood of a given text by multiplying the probability of each word in the text conditional on the previous one.

One difficulty is the potential for double-use of data: the first pass through a Markov chain account is already applying to the data a highly flexible Bayesian neural network with billions of parameters (one’s brain). If one spots an anomalous dataset and subsequent analysis confirms it, what does this mean? I am reminded of one past incident: someone had lost a great deal of money on a Bitcoin gambling website, and suspected the site had defrauded him. But he had contacted me only because he had had unusual losses. What does an analysis mean? Imagine that the top 1% of losers get angry and start looking into whether they were cheated; they go to a statistician who duly computes that based on the number of games played, there is a p=0.01 that they would lose as much or more as they did… If one had all the gambling records, one could look at the overall patterns and see if there are more losers than there should be given the rules of the game and a supposedly fair random number generator, but what does one do with 1 self-selected player? The data generation process is certainly neither random nor ignorable nor modelable without dubious assumptions.

A few possible attacks come to mind:

• observation of malformed syntax or lack of long-range dependencies
• vocabulary or output outside an independently trained Markov chain’s domain
• unusually low likelihood for an independently trained Markov chain to generate known samples
• unusually low likelihood for an independently trained Markov chain to generate known samples compared to newly generated samples filtered at a 1-in-100s quality level
• unusually high quality of known samples compared to newly generated samples from independently trained Markov chain filtered at a 1-in-100s quality level, tested nonparametrically or parametrically as a mixture model

Markov chains produce realistic-looking output and are efficient to create & run, but, compared to RNNs, they notoriously model recursive syntax poorly, such as nested parentheses (since they have no way of remembering whether a parenthetical comment had been started), and cannot extrapolate - for example, a word-level Markov chain can’t create new words, and would require n-grams to have available fragments of words which could be recombined. The memory-less nature of Markov chains also means that, lacking any memory which could model the long-range correlations found in natural English text like systematic use of particular names/topics/vocabulary, larger samples quickly veer off-topic and become gibberish and lack any coherency possibly even inside a single sentence. (RNNs also have this problem, but somewhat less.)

With the limits of a Markov chain in mind, it would be easy to detect faked Markov chain output with large samples: it is just difficult for a human to deliberately generate long text which is as nonsensical and syntactically invalid as a Markov chain creates, for the same reason an unpracticed human is a remarkably bad random number generator. However, for this same reason the selected Markov samples tend to be very short, usually no more than a sentence. It might be possible to measure this on the samples as a whole and observe higher entropy or memoryless-ness (eg by measuring compression performance or efficiency of a Markov chain in modeling the samples), but I would guess that usually the samples are not long enough or large enough for this to have reasonable statistical power as a test. This eliminates the easiest test.

Since the corpus is known in many of these cases, we can assume access to a Markov chain model which is similar (if not identical) to the one which supposedly wrote all the tweets. This gives us several possibilities.

We could exploit the lack of creativity of Markov chains and look for anything in the tweets which is not present in the original corpus. For example, if a word like cromulent appears neither in the Puppet documentation nor (having been coined in 1996, 59 years after he died) in H.P. Lovecraft’s fiction, then it would have a probability of 0 of being generated by any Puppet/Lovecraft Markov chain (as no word will have any transition probability to it). Such extra-corporal vocabulary immediately proves human authorship.

Continuing this same logic, we could take the corpus, train our own Markov chain (which will at least be similar), and use it to calculate the likelihood of all the tweets. A human-written tweet may be possible for the Markov chain to have written, but it will be far more unlikely than most of the ones the Markov chain actually wrote & were selected. So we would see that most of the tweets have reasonable log likelihoods, but that our suspicious ones will be far more extreme. (If the Markov chains are word-level, this test subsumes the impossible-word test: any tweet with a word not in the corpus, and hence not represented in the Markov chain, will have a meaningless likelihood.)

This likelihood test might not help if they are all equally extreme, in which case one could use our Markov chain in another manner, as a generative model, to try to estimate the likelihood of getting as great a tweet. For this, one samples several thousand samples from our Markov chain, and screens them for good ones. This creates an empirical distribution of the likelihoods of good tweets conditional on the null hypothesis of a Markov chain author; in this case, the null hypothesis is known to be true by construction. Then to test, one compares the known-Markov-chain tweets with the likelihoods of the suspect tweets (perhaps with a permutation test). They should be similar.

Alternately, if one doesn’t want to use likelihoods as a measure of improbability, one could instead use some human measure of funniness like having rating the originals and the samples on a scale 1-5, and comparing them. The original poster is probably not screening more than a few hundred generated tweets for each selected tweet, so given a similar level of stringency, one’s generated tweets should be equally good; if the originals turn out to be extremely better than yours, to a level where you would have to screen thousands of random samples, that is highly suspicious and suggests the originals were too good to be true.

With ratings or likelihoods, one could try to assume a decreasing distribution like an exponential: most samples will be incoherent and totally unfunny, many will be slightly funny, a few will be funny, and a very few will be very funny. The ratings on samples generated from our Markov chain will probably follow a smooth distribution. However, if a human is authoring some in an attempt to spice things up, they will be above the average of the Markov chain (otherwise why bother with cheating?), and if there is a substantial number of them, this will create an anomaly in the ratings of the originals - a bump indicating that the tweets are coming from two different populations. In this case, it can be modeled as a mixture model with either k=1 or k=2, and the p-value or Bayesian posterior probability calculated for 1 vs 2.

Optimal Existential Risk Reduction Investment

An existential risk is any risk which destroys or permanently cripples human civilization, such as an asteroid strike or pandemic. Since humanity might otherwise continue for millions of years, creating untold trillions of humans and colonizing the galaxy, human extinction represents the loss of literally astronomical amounts of utility. The loss is greater than any disaster up to extinction levels, as humanity can always recover from lesser disasters; but there is no recovery from a total destruction. Thus, the expected value of even a slight reduction in an exotic risk ought to itself be astronomical, or at least extremely large; under plausible values for well-characterized x-risks like asteroid strikes or nuclear war or pandemic, preventing them may be the charitable spending with the highest expected value and they should be receiving all charitable expenditures.

This strikes people as odd and dangerous reasoning. Is it really true that we should be spending almost unlimited amounts of money on these things and not otherwise extremely compelling charities like distributing malaria nets in Africa to save millions of lives or vaccine distribution or funding research into ending aging? And if we should, how do we choose what fraction to spend on global warming rather than artificial intelligence? What if someone discovers an entirely new x-risk not previously considered, like nearby supernovas or vacuum collapses or nanotechnology grey goo?

Thinking historically, it’s clear in retrospect that someone concerned about x-risk would be better off not going after the terminal goal of x-risk reduction but instead spending their money on instrumental goals such as science/technology or economic growth.

Imagine someone in England in 1500 who reasons the same way about x-risk: humanity might be destroyed, so preventing that is the most important task possible. He then spends the rest of his life researching the Devil and the Apocalypse. Such research is, unfortunately, of no value whatsoever unless it produces arguments for atheism demonstrating that that entire line of enquiry is useless and should not be pursued further. But as the Industrial and Scientific Revolutions were just beginning, with exponential increases in global wealth and science and technology and population, ultimately leading to vaccine technology, rockets and space programs, and enough wealth to fund all manner of investments in x-risk reduction, he could instead had made a perhaps small but real contribution by contributing to economic growth by work & investment or making scientific discoveries.

For example, Isaac Newton’s discoveries in astronomy and the laws of motion helped inaugurate threads of work that led directly to space satellites which can watch for asteroids with Earth-crossing orbits. Isaac Newton himself was concerned with x-risk, as he feared that Halley’s Comet would, centuries hence, plunge into the Sun and cause expansion destroying the Earth and humanity. What could Newton have done to directly reduce this x-risk at the time? Absolutely nothing. There were no feasible counter-measures nor any foreseeable technologies which could forestall a comet or protect humanity from the Sun engulfing the Earth; there was not and still is not a mine or bomb shelter deep enough for that. What he could have done is close to what he did do: make fundamental advances in science which posterity could build on and one day be rich and wise enough to do something about the x-risk. As it happens, Newton was not quite right about Halley’s Comet (comets are not a meaningful fraction of the Sun’s mass) but there was an x-risk he was unaware of: giant asteroid impacts. And the solutions to a giant comet - observe all comets carefully to project their future orbits, destroy it, redirect its orbit, evacuate human colonists to safety to unaffected planets (Newton suggested the satellites of the gas giants) - are much the same as for a giant asteroid impact, and all benefit from economic growth & greater science/technology (someone has to pay for, and design those satellites and spacecraft, after all).

Economic wealth & science/technology are all-purpose goods: they are useful for compound growth, and can also be spent on x-risk reduction. They are the ultimate instrumental goods. If one is badly ignorant, or poor, or unable to meaningfully reduce an x-risk, one is better off accepting the x-risk and instead spending resources on fixing the former problems. One would prefer to get rid of the x-risk as soon as possible, of course, but given one’s starting position, there may simply be no better strategy and the risk must be accepted.

This raises the question: what is the optimal distribution of resources to economic growth vs x-risk reduction over time which maximizes expected utility?

Intuitively, we might expect something like early on investing nothing at all in x-risk reduction as there’s not much money available to be spent, and money spent now costs a lot of money down the line in lost compound growth; and then as the economy reaches modern levels and the opportunity cost of x-risk becomes dire, money is increasingly diverted to x-risk reduction. One might analogize it to insurance - poor people skimp on insurance because they need the money for other things which hopefully will pay off later like education or starting a business, while rich people want to buy lots of insurance because they already have enough and they fear the risks. If this were an investment question, a good strategy would be something like the Kelly criterion or probability matching strategies like Thompson sampling: even if the expected value of x-risk reduction is higher than other investments, it only pays off very rarely and so receives a very small fraction of one’s investments. However, it’s not clear that the Kelly criterion or Thompson sampling are optimal or even relevant: because while Kelly avoids bankruptcy in the form of gambler’s ruin but does so only by making arbitrarily small bets to avoid going bankrupt & refusing to ever risk one’s entire wealth; with x-risks, the bankruptcy (extinction) can’t be avoided so easily, as the risk is there whether you like it or not, and one cannot turn it to 0. (This comes up often in discussion of why the Kelly criterion is relevant to decision-making under risk; see also Peters 2011 and the niche area of evolutionary finance like Evstigneev et al 2008/Lensberg & Schenk-Hoppé 2006 which draws connections between the Kelly criterion, probability matching, long-term survival & evolutionary fitness.) In economics, similar questions are often dealt with in terms of the life-cycle hypothesis in which economic agents strive to maximize their utility over a career/lifetime while avoiding inefficient intertemporal allocation of wealth (as Mark Twain put it, when in youth a dollar would bring a hundred pleasures, you can’t have it. When you are old, you get it & there is nothing worth buying with it then. It’s an epitome of life. The first half of it consists of the capacity to enjoy without the chance; the last half consists of the chance without the capacity.); in the life-cycle, one tries to build wealth as quickly as possible while young, even going into debt for investments like a college education, then begins saving up, consuming some, until retirement, at which point one consumes it all until one dies. But as far as I’ve seen any results, life-cycle models tend to not include any mechanism for spending in order to reduce mortality/aging and accept the risk of death as a given.

We could create a simple Markov decision process model. An agent (humanity), each time period (year), has a certain amount of wealth and an x-risk probability P. In this period, it can choose to allocate that wealth between economic growth, in which case it receives that investment plus a return, and it can buy a permanent percentage reduction in the x-risk for a fixed sum. For the reward, the x-risk is binary sampled with probability P; if the sample is true, then the reward is 0 and the decision process terminates, else the reward is the wealth and the process continues. Let’s imagine that this process can run up to 10,000 time periods, with a starting wealth of $248 billion (Angus Deaton’s estimate of PPP world GDP in 1500 https://en.wikipedia.org/wiki/List_of_regions_by_past_GDP_%28PPP%29 ), the economic growth rate is 2% (the long-run real growth rate of the global economy), the existential risk probability is 0.1% per year (arbitrarily chosen), and one can buy a reduction of 1% for a billion dollars. (We’ll work in trillions units to help numeric stability.) What strategy maximizes the cumulative rewards? A few simple ones come to mind: 1. the agent could simply ignore the x-risk and reinvests all wealth, which to a first approximation, is the strategy which has been followed throughout human history and is primarily followed now (lumping together NASA’s Spaceguard program, biowarfare and pandemic research, AI risk research etc probably doesn’t come to more than$1-2b a year in 2016). This maximizes economic growth rate but may backfire as the x-risk never gets reduced.
2. the agent could spend the full gain in its wealth from economic growth (2%) on x-risk reduction. The wealth doesn’t grow and the returns from x-risk reduction do diminish, but the x-risk is at least reduced greatly over time.
3. the agent could implement a sort of probability matching: it spends on x-risk reduction a fraction of its wealth equal to the current P. This reduces how much is spent on extremely small x-risk reductions, but it might be suboptimal because it’ll pay the largest fraction of its economy in the first time period, then second-largest in the second time period and so on, losing out on the potential compounding.
4. a more complicated hybrid strategy might work: it maximizes wealth like #1 for the first n time periods (eg n=516), and then it switches to #2 for the remaining time period
5. like #4, but switching from #1 to #3 for the remaining time periods.
constantInvestmentAgent <- function (t, w, xrp) { return(c(w, 0)) }
constantReductionAgent  <- function (t, w, xrp) { drawdown <- 0.9803921573; return(c(drawdown*w, (1-drawdown)*w)) }
probabilityMatchAgent   <- function (t, w, xrp) { return(c(w*(1-xrp), w*xrp)) }
investThenReduceAgent   <- function (t, w, xrp, n=516) { if (t<n) { return(constantInvestmentAgent(t, w, xrp)) } else { return(constantReductionAgent(t, w, xrp)) } }
investThenMatchAgent    <- function (t, w, xrp, n=516) { if (t<n) { return(constantInvestmentAgent(t, w, xrp)) } else { return(probabilityMatchAgent(t, w, xrp)) } }

simulateWorld <- function(agent, t=10000) {
initialW <- 0.248
initialP <- 0.001
df <- data.frame(T=0, Wealth=initialW, XriskP=initialP)

for (i in 1:t) {
last <- tail(df, n=1)
xrisk <- rbinom(1,1, p=last$XriskP) if (xrisk) { break; } else { choices <- agent(last$T, last$Wealth, last$XriskP)
newXriskP <- last$XriskP * (1 - 0.01)^(choices[2] / 0.001) newWealth <- choices[1] * 1.02 df <- rbind(df, data.frame(T=i, Wealth=newWealth, XriskP=newXriskP)) } } df$Reward <- cumsum(df$Wealth) return(df) } library(parallel); library(plyr) simulateWorlds <- function(agent, iters=1000) { mean(ldply(mclapply(1:iters, function(i) { tail(simulateWorld(agent), n=1)$Reward }))$V1) } simulateWorlds(constantReductionAgent) # [1] 2423.308636 simulateWorlds(investThenReduceAgent) # [1] 10127204.73 simulateWorlds(constantInvestmentAgent) # [1] 1.154991741e+76 simulateWorlds(investThenMatchAgent) # [1] 7.53514145e+86 ## Optimize the switch point: which.max(sapply(seq(1, 10000, by=100), function(N) { simulateWorlds(function(t,w,xrp) { investThenMatchAgent(t, w, xrp, n=N) }, iters=100)})) # [1] 3 simulateWorlds(function(t,w,xrp) { investThenMatchAgent(t, w, xrp, n=300) }) # [1] 9.331170221e+86 simulateWorlds(probabilityMatchAgent) # [1] 1.006834082e+87 So of our 5 strategies, the constant reduction agent performs the worst (probably because with economic growth choked off, it can only buy small x-risk reductions), followed by the invest-then-reduce agent; then the get rich before you get old constant investment agent manages to often attain very high growth rates when it’s lucky enough that x-risks don’t strike early on; but far better than any of them, by orders of magnitude, are the partial and full probability matching agents. The partial probability matching agent turns out to have a suboptimal switch point t=516, and a more careful search of switch points finds that t~=300 is the best switch point and it exceeds the pure probability matcher which matches from the start. What’s going on there? I suspect it’s something similar to the difference in multi-armed bandit problems between the asymptotically optimal solution and the optimal solution for a fixed horizon found using dynamic programming: in the former scenario, there’s an indefinite amount of time to do any exploration or investment in information, but in the latter, there’s only a finite time left and exploration/growth must be done up front and then the optimal decision increasingly shifts to exploitation rather than growth. Why does probability matching in general work so well? It may simply be because it’s the only baseline strategy which adjusts its xrisk investment over time. This doesn’t demonstrate that probability matching is optimal, just that it beats the other baseline strategies. Other strategies could be used to decrease xrisk investment over time - instead of being proportional to xrisk P, it could shrink linearly over time, or by square root, or logarithmically, or… What reinforcement learning techniques might we use to solve this? This problem represents a large Markov Decision Process with 1 discrete state variable (time, t=0-10000), 2 continuous state variables (wealth, and risk probability), and 1 continuous action (fraction of growth to allocate to the economy vs existential risk reduction). The continuous action can be discretized into 11 actions without probably losing anything (allocate 100%/90%..10%/0%), but the 2 state variables can’t be discretized easily because they can span many orders of magnitude. • dynamic programming a decision tree with backwards induction: optimal, but requires discrete actions and state variables, and even if discretized, 10000 time steps would be infeasibly large. • standard tabular learning: Q-learning, SARSA, temporal differences: requires discrete actions and state variables • Deep Q-Networks: requires discrete actions, but not state variables • MDP solvers: value iteration etc: optimal, but requires discrete actions and state variables • hybrid MDP solvers: optimal, and can handle a limited amount of continuous state variables (but not continuous actions), which would work here; but high quality software implementations are rarely available. One such hybrid MDP solver is hmpd, which solves problems specified in the PDDL Lisp-like DSL (judging from the examples, a version with probabilistic effects, so PPDDL 1.0?). After trying to write down a PPDDL model corresponding to this scenario, it seems that PPDDL is unable to represent probabilities or rewards which change with time and so cannot represent the increase in wealth or decrease in x-risk probability. • policy gradients: can handle continuous state variables & actions but are highly complex and unstable; high quality software implementations are unavailable Of the possible options, a DQN agent seems like the best choice: a small neural network should be able to handle the problem and DQN only requires the actions to be discretized. reinforce.js provides a DQN implementation in JS which I’ve used before, so I start there by rewriting the problem in JS var script = document.createElement("script"); script.src = "https://www.gwern.net/docs/rl/armstrong-controlproblem/2016-02-02-karpathy-rl.js"; document.body.appendChild(script); // environment: t, w, xrp function simulate(environment, w_weight, xrp_weight) { var xrisk = Math.random() < environment.xrp if (xrisk) { return {reward: -100, alive: false, t: environment.t, w: environment.w, xrp: environment.xrp}; } else { return {reward: Math.log(environment.w), alive: true, t: environment.t+1, w: environment.w*w_weight*1.02, xrp: environment.xrp * (Math.pow((1 - 0.01), (xrp_weight / 0.001))) } } } var defaultState = {t: 0, w: 0.248, xrp: 0.01} // simulate(defaultState, 0.99, 0.01) // simulate(defaultState, 0.99, 0.01) var env = {}; env.getNumStates = function() { return 3; }; // there are only 3 state variables: t/w/xrp env.getMaxNumActions = function() { return 11; }; // we'll specify 10 possible allocations: 1/0, 0.998/0.002 .. 0.98/0.02 var spec = { num_hidden_units: 200, experience_add_every: 20, learning_steps_per_iteration: 1, experience_size: 1000000, alpha: 0.01, epsilon: 1.0, gamma: 0.99 // minimal discounting }; var agent = new RL.DQNAgent(env, spec); var total_reward = 0; state = defaultState; spec.epsilon = 1.0; // reset epsilon if we've been running the loop multiple times for(var i=0; i < 10000*3000; i++) { var action = agent.act(state) state = simulate(state, 1-(action/500), 0+(action/500) ); agent.learn(state.reward); total_reward = total_reward + state.reward; if (Number.isInteger(Math.log(i) / Math.log(10)) ) { spec.epsilon = spec.epsilon / 1.5; } // decrease exploration if (!state.alive || state.t >= 10000) { // if killed by x-risk or horizon reached console.log(state.t, state.w, state.xrp, total_reward); total_reward = 0; state = defaultState; } } //exercise the trained agent to see how it thinks total_reward=0 state=defaultState; spec.epsilon = 0; for (var t=0; t < 10000; t++) { action = agent.act(state) state = simulate(state, 1-(action/500), 0+(action/500) ); total_reward = total_reward + state.reward console.log(action, state, total_reward); } After a day of training, the DQN agent had learned to get up to 5e41, which was disappointingly inferior to the constant investment & probability matching agents (1e87). The NN looks big enough for this problem and the experience replay buffer was more than adequate; NNs in RL are known to have issues with the reward, though, and typically clamp the reward to a narrow range, so I suspected that rewards going up to 5e41 (interpreting wealth on each turn as the reward) might be playing havoc with convergence, and switched the reward to log wealth instead. This did not make a noticeable difference overnight (aside from the DQN agent now achieving 9.5e41). I wondered if the risk was too rare for easy learning and 100 neurons was not enough to approximate the curve over time, so I fixed a bug I noticed where the simulation did not terminate at t=10000, doubled led the neuron count, increased the initial x-risk to 1%, and began a fresh run. After 1 day, it reached 9.4e41 total reward (unlogged). Cumulative log score for DQN after tweaks and ~2h of training: regularly reaches ~470k when it doesn’t die immediately (which happens ~1/20 of the time). In comparison, probability-matching agent averages a cumulative log score of 866k. After 2 days of training, the DQN had improved only slightly; the on-policy strategy appears mostly random aside from having driven the xrisk probability down to what appears to be the smallest float JS supports, so it still had not learned a meaningful compromise between xrisk reduction and investment. TODO: revisit with MCTS at some point? Model Criticism via Machine Learning Here’s how we put it on the very first page of our book: The process of Bayesian data analysis can be idealized by dividing it into the following three steps: 1. Setting up a full probability model - a joint probability distribution for all observable and unobservable quantities in a problem. The model should be consistent with knowledge about the underlying scientific problem and the data collection process. 2. Conditioning on observed data: calculating and interpreting the appropriate posterior distribution - the conditional probability distribution of the unobserved quantities of ultimate interest, given the observed data. 3. Evaluating the fit of the model and the implications of the resulting posterior distribution: how well does the model fit the data, are the substantive conclusions reasonable, and how sensitive are the results to the modeling assumptions in step 1? In response, one can alter or expand the model and repeat the three steps. How does this fit in with goals of performing statistical analysis using artificial intelligence? 3. The third step - identifying model misfit and, in response, figuring out how to improve the model - seems like the toughest part to automate. We often learn of model problems through open-ended exploratory data analysis, where we look at data to find unexpected patterns and compare inferences to our vast stores of statistical experience and subject-matter knowledge. Indeed, one of my main pieces of advice to statisticians is to integrate that knowledge into statistical analysis, both in the form of formal prior distributions and in a willingness to carefully interrogate the implications of fitted models. One way of looking at step #3 is to treat the human statistician as another model: specifically, he is a large neural network with trillions of parameters, who has been trained to look for anomalies & model misspecification, and to fix them when he finds them, retraining the model, until he can no longer easily distinguish the original data from the model’s predictions or samples. As he is such a large model with the ability to represent and infer a large class of nonlinearities, he can usually easily spot flaws where the current model’s distribution differs from the true distribution. This bears a considerable resemblance to the increasing popularity of generative adversarial networks (GANs): using pairs of neural networks, one of which tries to generate realistic data, and a second which tries to classify or discriminate between real and realistic data. As the second learns ways in which the current realistic data is unrealistic, the first gets feedback on what it’s doing wrong and fixes it. So the loop is very similar, but fully automated. (A third set of approaches this resembles is actor-critic reinforcement learning algorithms.) If we consider the kinds of models which are being critiqued, and what is critiquing, this gives us 4 possible combinations: simple complex simple model fit indexes+linear model statistician+linear model complex model fit indexes+ML ML+ML (eg GANs) 1. Simple/simple is useful for cases like linear regression where classic methods like examining residuals or R^2s or Cook indexes can often flag problems with the model. 2. Simple/complex is also useful, as the human statistician can spot additional problems. 3. Complex/simple is probably useless, as the NNs may easily have severe problems but will have fit any simple linear structure and fool regular diagnostics. 4. Complex/complex can be very useful in machine learning, but in different ways from a good simple model. Fast, simple, general - a good statistical method lets you choose one; a great method lets you choose two. (Consider linear models, decision trees, NNs, MCMC, ABC, discrete Bayesian networks, and exponential family vs nonparametric methods as examples of the tradeoffs here.) So is quadrant 2 fully populated by human statisticians? We wouldn’t necessarily want to use GANs for everything we use statisticians for now, because neural networks can be too powerful and what we want from our models is often some sort of clear answer like does X predict Y? and simplicity. But we could replace the statistician with some other powerful critic from machine learning - like a NN, SVM, random forest, or other ensemble. So instead of having two NNs fighting each other as in a GAN, we simply have one specified model, and a NN which tries to find flaws in it, which can then be reported to the user. The loop then becomes: write down and fit a model to the real data; generative posterior predictive samples from the distribution; train a small NN on real data vs predictive data; the classification performance measures the plausibility of the predictive samples (perhaps something like a KL divergence), giving a measure of the model quality, and flags data points which are particularly easily distinguished as real; the human statistician now knows exactly which data points are not captured by the model and can modify the model; repeat until the NN’s performance declines to chance. Let’s try an example. We’ll set up a simple linear model regression Y ~ A + B + C with a few problems in it: 1. the trend is not linear but slightly quadratic 2. the outcome variable is also right-censored at a certain point 3. and finally, the measured covariates have been rounded set.seed(2016-11-23) n <- 10000 ceiling <- 1 a <- rnorm(n) b <- rnorm(n) c <- rnorm(n) y <- 0 + 0.5*a + 0.5*b + 0.5*c^2 + rnorm(n) y_censored <- ifelse(y>=3, 3, y) df <- data.frame(Y=y_censored, A=round(a, digits=1), B=round(b, digits=1), C=round(c, digits=1)) l <- lm(Y ~ A + B + C, data=df) summary(l) plot(l) plot(df$Y, predict(l, df))

l2 <- lm(Y ~ A + B + I(C^2), data=df)
summary(l2)
plot(df$Y, predict(l2, df)) The censoring shows up immediately on the diagnostics as an excess of actual points at 3, but the quadraticity is subtler, and I’m not sure I can see the rounding at all. library(randomForest) ## First, random forest performance under the null hypothesis modelNull <- data.frame(Y=c(df$Y, df$Y), Real=c(rep(1, n), rep(0, n)), A=c(df$A, df$A), B=c(df$B, df$B), C=c(df$C, df$C)) r_n <- randomForest(as.ordered(Real) ~ Y + A + B + C, modelNull); r_n # Type of random forest: classification # Number of trees: 500 # No. of variables tried at each split: 2 # # OOB estimate of error rate: 100% # Confusion matrix: # 0 1 class.error # 0 0 10000 1 # 1 10000 0 1 modelPredictions <- data.frame(Y=c(df$Y, predict(l, df)), Real=c(rep(1, n), rep(0, n)), A=c(df$A, df$A), B=c(df$B, df$B), C=c(df$C, df$C))
r <- randomForest(as.ordered(Real) ~ Y + A + B + C, modelPredictions); r
#                Type of random forest: classification
#                      Number of trees: 500
# No. of variables tried at each split: 2
#
#         OOB estimate of  error rate: 6.59%
# Confusion matrix:
#      0    1 class.error
# 0 9883  117      0.0117
# 1 1200 8800      0.1200

## many of the LM predictions are identical, but the RF is not simply memorizing them as we can jitter predictions and still get the same classification performance:
modelPredictions$Y2 <- jitter(modelPredictions$Y)
randomForest(as.ordered(Real) ~ Y2 + A + B + C, modelPredictions)
#...                Type of random forest: classification
#                      Number of trees: 500
# No. of variables tried at each split: 2
#
#         OOB estimate of  error rate: 6.57%
# Confusion matrix:
#      0    1 class.error
# 0 9887  113      0.0113
# 1 1200 8800      0.1200

Note we need to be careful about collecting the posterior predictive samples: if we collect 10000 posterior samples for each of the 10000 datapoints, we’ll store 100002 numbers which may cause problems. 1 should be enough.

library(runjags)
model <- 'model {
for (i in 1:n) {
mean[i] <- mu + betaA*A[i] + betaB*B[i] + betaC*C[i]
Y[i] ~ dnorm(mean[i], tau)
}

sd   ~ dgamma(0.01, 0.01)
tau  <- 1/sqrt(sd)

mu ~ dnorm(0, 100)
betaA ~ dnorm(0, 100)
betaB ~ dnorm(0, 100)
betaC ~ dnorm(0, 100)
}'
model <- run.jags(model, data = with(df, list(Y=c(Y, rep(NA, nrow(df))), A=c(A,A), B=c(B,B), C=c(C,C), n=2*nrow(df))), inits=list(mu=0.45, sd=0.94, betaA=0.47, betaB=0.46, betaC=0), monitor=c("Y"), n.chains = 1, sample=1)

posterior_predictive <- tail(n=10000, model$mcmc[[1]][1,]) plot(df$Y, posterior_predictive)

modelPredictions_r <- data.frame(Y=c(df$Y, posterior_predictive), Real=c(rep(1, n), rep(0, n)), A=c(df$A, df$A), B=c(df$B, df$B), C=c(df$C, df$C)) r <- randomForest(as.ordered(Real) ~ Y + A + B + C, modelPredictions_r); r # OOB estimate of error rate: 49.11% # Confusion matrix: # 0 1 class.error # 0 4953 5047 0.5047 # 1 4776 5224 0.4776 model_rounded <- 'model { for (i in 1:n) { roundA[i] ~ dround(A[i], 3) roundB[i] ~ dround(B[i], 3) roundC[i] ~ dround(C[i], 3) mean[i] <- mu + betaA*roundA[i] + betaB*roundB[i] + betaC*roundC[i] Y[i] ~ dnorm(mean[i], tau) } sd ~ dgamma(0.01, 0.01) tau <- 1/sqrt(sd) mu ~ dnorm(0, 100) betaA ~ dnorm(0, 100) betaB ~ dnorm(0, 100) betaC ~ dnorm(0, 100) }' model_r <- run.jags(model_rounded, data = with(df, list(Y=c(Y, rep(NA, nrow(df))), A=c(A,A), B=c(B,B), C=c(C,C), n=2*nrow(df))), inits=list(mu=0.45, sd=0.94, betaA=0.47, betaB=0.46, betaC=0), monitor=c("Y"), n.chains = 1, sample=1) posterior_samples <- tail(n=10000, model_r$mcmc[[1]][1,])
posterior_predictive <- ifelse(posterior_samples>=3, 3, posterior_samples)
plot(df$Y, posterior_predictive) modelPredictions_r <- data.frame(Y=c(df$Y, posterior_predictive), Real=c(rep(1, n), rep(0, n)), A=c(df$A, df$A), B=c(df$B, df$B), C=c(df$C, df$C))
r_r <- randomForest(as.ordered(Real) ~ Y + A + B + C, modelPredictions_r); r_r
#         OOB estimate of  error rate: 50.48%
# Confusion matrix:
#      0    1 class.error
# 0 4814 5186      0.5186
# 1 4909 5091      0.4909
model_rounded_censor <- 'model {
for (i in 1:n) {
roundA[i] ~ dround(A[i], 3)
roundB[i] ~ dround(B[i], 3)
roundC[i] ~ dround(C[i], 3)
mean[i] <- mu + betaA*roundA[i] + betaB*roundB[i] + betaC*roundC[i]
Y[i] ~ dnorm(mean[i], tau)
is.censored[i] ~ dinterval(Y[i], c)
}

sd   ~ dgamma(0.01, 0.01)
tau  <- 1/sqrt(sd)

mu ~ dnorm(0, 100)
betaA ~ dnorm(0, 100)
betaB ~ dnorm(0, 100)
betaC ~ dnorm(0, 100)
}'
model_r_c <- run.jags(model_rounded_censor, data = with(df, list(Y=c(Y, rep(NA, nrow(df))), A=c(A,A), B=c(B,B), C=c(C,C), n=2*nrow(df), is.censored=c(as.integer(Y==3), as.integer(Y==3)), c=3)), inits=list(mu=0.37, sd=1, betaA=0.42, betaB=0.40, betaC=0), monitor=c("Y"), n.chains = 1, adapt=0, burnin=500, sample=1)

posterior_samples <- tail(n=10000, model_r_c$mcmc[[1]][1,]) posterior_predictive <- ifelse(posterior_samples>=3, 3, posterior_samples) modelPredictions_r_c <- data.frame(Y=c(df$Y, posterior_predictive), Real=c(rep(1, n), rep(0, n)), A=c(df$A, df$A), B=c(df$B, df$B), C=c(df$C, df$C))
r_r_c <- randomForest(as.ordered(Real) ~ Y + A + B + C, modelPredictions_r_c); r_r_c
#         OOB estimate of  error rate: 53.67%
# Confusion matrix:
#      0    1 class.error
# 0 4490 5510      0.5510
# 1 5224 4776      0.5224
model_rounded_censor_quadratic <- 'model {
for (i in 1:n) {
roundA[i] ~ dround(A[i], 3)
roundB[i] ~ dround(B[i], 3)
roundC[i] ~ dround(C[i], 3)
mean[i] <- mu + betaA*roundA[i] + betaB*roundB[i] + betaC*roundC[i]^2
Y[i] ~ dnorm(mean[i], tau)
is.censored[i] ~ dinterval(Y[i], c)
}

sd   ~ dgamma(0.01, 0.01)
tau  <- 1/sqrt(sd)

mu ~ dnorm(0, 100)
betaA ~ dnorm(0, 100)
betaB ~ dnorm(0, 100)
betaC ~ dnorm(0, 100)
}'

model_r_c_q <- run.jags(model_rounded_censor_quadratic, data = with(df, list(Y=c(Y, rep(NA, nrow(df))), A=c(A,A), B=c(B,B), C=c(C,C), n=2*nrow(df), is.censored=c(as.integer(Y==3), as.integer(Y==3)), c=3)), inits=list(mu=0.37, sd=1, betaA=0.42, betaB=0.40, betaC=0), monitor=c("Y"), n.chains = 1, adapt=0, burnin=500, sample=1)

posterior_samples <- tail(n=10000, model_r_c_q$mcmc[[1]][1,]) posterior_predictive <- ifelse(posterior_samples>=3, 3, posterior_samples) modelPredictions_r_c_q <- data.frame(Y=c(df$Y, posterior_predictive), Real=c(rep(1, n), rep(0, n)), A=c(df$A, df$A), B=c(df$B, df$B), C=c(df$C, df$C))
r_r_c_q <- randomForest(as.ordered(Real) ~ Y + A + B + C, modelPredictions_r_c_q); r_r_c_q
#         OOB estimate of  error rate: 61.02%
# Confusion matrix:
#      0    1 class.error
# 0 3924 6076      0.6076
# 1 6127 3873      0.6127

trueNegatives <- modelPredictions_r_c_q[predict(r_r_c_q) == 0 & modelPredictions_r_c_q$Real == 0,] Where can we go with this? The ML techniques can be used to rank existing Bayesian models in an effective if unprincipled way. Techniques which quantify uncertainty like Bayesian neural networks could give more effective feedback by highlighting the points the Bayesian NN is most certain are fake, guiding the analyst towards the worst-modeled datapoints and providing hints for improvement. More inspiration could be borrowed from the GAN literature, such as minibatch discrimination - as demonstrated above, the random forests only see one data point at a time, but in training GANs, it has proven useful to instead examine multiple datapoints at a time to encourage the generator to learn how to generate a wide variety of datapoints rather than modeling a few datapoints extremely well; a ML model which can predict multiple outputs simultaneously based on multiple inputs would be analogous (that is, instead of X ~ A + B + C, it would look more like X1 + X2 + X3 ... ~ A1 + B1 + C1 + A2 + B2 + C2 + ..., with the independent & dependent variables from multiple data points all fed in simultaneously as a single sample) and might be an even more effective model critic. Proportion of Important Thinkers by Global Region Over Time in Charles Murray’s Human Accomplishment Human Accomplishment is a 2003 book by Charles Murray reporting a large-scale citation analysis of biographical dictionaries & reference books on art/literature/science/mathematics/philosophy/science throughout history, quantifying the relative importance of significant individuals such as Isaac Newton or Immanuel Kant or Confucius and the temporal & geographical patterns; in particular, it demonstrates large European contributions throughout history and increasingly dramatically post-1400 AD. The dataset has been released. Emil Kirkegaard created a visualization of of the proportion by rough geographic region (European/Asian/other) in R using ggplot2 and LOESS smoothing. Perhaps the most striking aspect of it is the Dark Ages showing up as a spike in Asian proportion. This visualization could be improved in a few ways: • LOESS ignores the constraint that proportions must be 0-1 and naively extrapolates beyond the boundaries • no visualization of uncertainty is provided, either in the form of graphing the raw data points by superimposing a scatterplot or by providing standard errors or credible intervals • LOESS can provide estimate local standard errors & confidence intervals but they are of questionable meaning in the absence of the underlying counts • alternatively, the distribution of significant figures may not be treated correctly parametrically • proportions may reflect a time-series with trends and so precision is overstated These points can be addressed by: 1. switching from a LOESS plot to either splines or local binomial regressions 2. plotting the raw proportions grouped by decade or century 3. using a nonparametric bootstrap to calculate confidence intervals, a procedure which lends itself to visualization as an animation of plots of all the resamples, giving an intuitive sense of how important sampling error is to the overall pattern of curves and specific parts of history 4. alternately, instead of attempting to fit the proportion, one can fit the original count of significant figures in a binomial or log-normal Bayesian time-series model and sample from the posterior estimates of each region for each decade/century, and calculate posterior proportions, gaining full quantification of uncertainty, incorporation of any autocorrelation, and smoothing I didn’t realize Kirkegaard’s R code was available so I wound up redoing it myself (and getting the same results): ## export CSV from spreadsheet in https://osf.io/z9cnk/ h <- read.csv("HA.csv", header=TRUE) summary(h) # Serial Name Fl Birth Death Inventory ScienceField # Min. : 11.00 Descartes, René : 4 Min. :-700.000 Min. :-640.000 Min. :-559.00 Science :1442 :2560 # 1st Qu.: 6144.50 Hooke, Robert : 4 1st Qu.:1557.250 1st Qu.:1580.000 1st Qu.:1638.00 Lit.West : 835 Tech : 239 # Median :12534.50 Leonardo da Vinci: 4 Median :1804.000 Median :1782.000 Median :1844.00 Music.West: 522 Phys : 218 # Mean :15994.27 Archimedes : 3 Mean :1585.638 Mean :1616.174 Mean :1682.81 Art.West : 479 Chem : 204 # 3rd Qu.:21999.75 Bacon, Francis : 3 3rd Qu.:1900.000 3rd Qu.:1863.000 3rd Qu.:1930.00 Phil.West : 155 Biol : 193 # Max. :43134.00 d’Alembert, Jean : 3 Max. :1949.000 Max. :1910.000 Max. :1997.00 Art.China : 111 Math : 191 # (Other) :3981 NA's :304 NA's :351 (Other) : 458 (Other): 397 # Index Duplicate BirthCountry WorkCountry Ethnicity Woman No..of.Inventories # Min. : 0.60000 Min. :0.00000000 France : 564 France : 605 Germanic: 592 Min. :0.00000000 Min. :2.000000 # 1st Qu.: 3.54000 1st Qu.:0.00000000 Germany: 556 Britain: 574 French : 565 1st Qu.:0.00000000 1st Qu.:2.000000 # Median : 7.60000 Median :0.00000000 Britain: 554 Germany: 525 English : 441 Median :0.00000000 Median :2.000000 # Mean : 12.95713 Mean :0.06221889 Italy : 400 Italy : 406 Italian : 397 Mean :0.02198901 Mean :2.228916 # 3rd Qu.: 15.89000 3rd Qu.:0.00000000 USA : 306 USA : 375 USA : 276 3rd Qu.:0.00000000 3rd Qu.:2.000000 # Max. :100.00000 Max. :1.00000000 China : 239 China : 239 Chinese : 240 Max. :1.00000000 Max. :4.000000 # NA's :115 (Other):1383 (Other):1278 (Other) :1491 NA's :3753 levels(h$Ethnicity)
#  [1] "Ancient Greek" "Ancient Roman" "Arabic"        "Australian"    "Basque"        "Black"         "Bulgarian"     "Canadian"
#  [9] "Chinese"       "Croatian"      "Czech"         "Danish"        "Dutch"         "English"       "Estonian"      "Finnish"
# [17] "Flemish"       "French"        "Germanic"      "Greek"         "Hungarian"     "Icelandic"     "Indian"        "Irish"
# [25] "Italian"       "Japanese"      "Jewish"        "Latino"        "New Zealand"   "Norwegian"     "Polish"        "Portuguese"
# [33] "Romanian"      "Scots"         "Slavic"        "Slovenian"     "Spanish"       "Swedish"       "Swiss"         "USA"

european <- c("Ancient Greek", "Ancient Roman", "Australian", "Basque", "Bulgarian", "Canadian", "Croatian", "Czech", "Danish",
"Dutch", "English", "Estonian", "Finnish", "Flemish", "French", "Germanic", "Greek", "Hungarian", "Icelandic", "Irish",
"Italian", "Jewish", "New Zealand", "Norwegian", "Polish", "Portuguese", "Romanian", "Scots", "Slavic", "Slovenian",
"Spanish", "Swedish", "Swiss", "USA")
asian    <- c("Chinese", "Indian", "Japanese")
other    <- c("Arabic", "Black", "Latino")
groupMembership <- function(e) { if (e %in% european) { "European" } else { if (e %in% asian) { "Asian" } else { "Other" } } }
h$Group <- as.factor(sapply(h$Ethnicity, groupMembership))
summary(h$Group) # Asian European Other # 507 3379 116 ## We use 'Fl' (floruit/flourished), when a person is believed to have done their most important work, ## since birth/death is often unavailable. ## group to decades by rounding: h$Fl.decade <- round(h$Fl, digits=-1) hd <- subset(select=c(Fl.decade, Group), h) hdcount <- aggregate(cbind(Group) ~ Fl.decade+Group, length, data=hd) colnames(hdcount)[3] <- "Count" ## sort by time: hdcount <- hdcount[order(hdcount$Fl.decade),]
nrow(h); sum(hdcount$Count) # [1] 4002 # [1] 4002 head(hdcount, n=20) # Fl.decade Group Count # 178 -700 European 3 # 179 -680 European 1 # 180 -650 European 1 # 1 -600 Asian 2 # 181 -600 European 2 # 182 -580 European 2 # 183 -570 European 2 # 2 -550 Asian 1 # 184 -550 European 1 # 185 -540 European 5 # 3 -520 Asian 1 # 186 -520 European 3 # 4 -510 Asian 1 # 187 -510 European 2 # 188 -500 European 2 # 189 -480 European 6 # 190 -460 European 3 # 191 -450 European 7 # 5 -440 Asian 1 # 192 -440 European 11 ## One issue with the count data: decades with zero significant figures from a group ## (which happens frequently) get suppressed. Some tools can handle the omission ## automatically but many cannot, so we need to manually insert any missing decades with '0' decades <- seq(-700, 1950, by=10) for (i in 1:length(decades)) { d <- decades[i] if (nrow(hdcount[hdcount$Fl.decade==d & hdcount$Group=="European",])==0) { hdcount <- rbind(hdcount, data.frame(Fl.decade=d, Group="European", Count=0))} if (nrow(hdcount[hdcount$Fl.decade==d & hdcount$Group=="Asian",])==0) { hdcount <- rbind(hdcount, data.frame(Fl.decade=d, Group="Asian", Count=0))} if (nrow(hdcount[hdcount$Fl.decade==d & hdcount$Group=="Other",])==0) { hdcount <- rbind(hdcount, data.frame(Fl.decade=d, Group="Other", Count=0))} } hdcount <- hdcount[order(hdcount$Fl.decade),]

library(ggplot2); library(gridExtra)
c1 <- with(hdcount, qplot(Fl.decade, Count, color=Group) + stat_smooth())
c2 <- with(hdcount, qplot(Fl.decade, log1p(Count), color=Group) + stat_smooth())
grid.arrange(c1, c2, ncol=1)

The growth in human population and accomplishment post-1400 is so dramatic that it obscures earlier temporal variations:

Log-transformed, we can still see the inverted-V shape of European counts, but it’s somewhat subtle and does leave room for doubt about sampling error. Moving on to reproducing the proportions plot:

## Create proportions by summing per decade, then looping over each group & dividing by total for that decade:
for (i in 1:nrow(hdcount)) {
total <- decadeTotals[decadeTotals$Fl.decade == hdcount[i,]$Fl.decade,]$Count p <- hdcount[i,]$Count / total
hdcount$Proportion[i] <- if(is.nan(p)) { 0 } else { p } hdcount$Total[i] <- total
}
with(hdcount, qplot(Fl.decade, Proportion, color=Group) + stat_smooth() + coord_cartesian(ylim = c(0, 1)))

We successfully reproduce it, modulo the LOESS standard errors (which can be disabled by adding se=FALSE to stat_smooth()), including the unwanted nonsensical extrapolations. It is possible with some tricky ggplot2 functionality to add in binomial smoothing (along with some jitter to unbunch the datapoints at the modal 0).

## roughly equivalent to:
# glm(cbind(Count,Total) ~ splines::ns(Fl.decade,3), family="binomial", data=hdcount, subset=Group=="European")
binomial_smooth <- function(...) { geom_smooth(se=FALSE, method = "glm", method.args = list(family = "binomial"), ...) }
binomial_smooth(formula = y ~ splines::ns(x, 3)) +
geom_jitter(aes(color=Group), width=0.013,, height=0.013))

This still doesn’t provide any indication of sampling error uncertainty, however. Kirkegaard provides one with CIs derived from bootstrapping, so I will provide something a little different: visualizing the uncertainty dynamically by graphing the smoothed proportions for each resample in an animation of hundreds of bootstrap samples.

So to do this bootstrap, we package up the various transformations from before, so we can sample-with-replacement the original dataset14, transform, and plot repeatedly:

transformAndProportion <- function(df) {
df$Fl.decade <- round(df$Fl, digits=-1)
dfdcount <- aggregate(cbind(Group) ~ Fl.decade+Group, length, data=dfd)
colnames(dfdcount)[3] <- "Count"
if (nrow(dfdcount[dfdcount$Fl.decade==d & dfdcount$Group=="European",])==0) {
dfdcount <- rbind(dfdcount, data.frame(Fl.decade=d, Group="European", Count=0))}
if (nrow(dfdcount[dfdcount$Fl.decade==d & dfdcount$Group=="Asian",])==0) {
dfdcount <- rbind(dfdcount, data.frame(Fl.decade=d, Group="Asian", Count=0))}
if (nrow(dfdcount[dfdcount$Fl.decade==d & dfdcount$Group=="Other",])==0) {
dfdcount <- rbind(dfdcount, data.frame(Fl.decade=d, Group="Other", Count=0))}
}
dfdcount <- dfdcount[order(dfdcount$Fl.decade),] decadeTotals <- aggregate(Count ~ Fl.decade, sum, data=dfdcount) for (i in 1:nrow(dfdcount)) { p <- dfdcount[i,]$Count / decadeTotals[decadeTotals$Fl.decade == dfdcount[i,]$Fl.decade,]$Count dfdcount$Proportion[i] <- if(is.nan(p)) { 0 } else { p }
}
return(dfdcount)
}

bootPlot <- function(df) {
n <- nrow(df)
bootDf <- df[sample(1:n, n, replace=TRUE),]
bootDfdcount <- transformAndProportion(bootDf)
## WARNING: can't just call qplot due to animation/ggplot2 bug; have to assign & 'print'
p <- with(bootDfdcount, qplot(Fl.decade, Proportion, color=Group) +
binomial_smooth(formula = y ~ splines::ns(x, 3)) +
geom_jitter(aes(color=Group), width=0.013,, height=0.013))
print(p)
}
library(animation)
saveGIF({for (i in 1:200) { bootPlot(h) }}, interval=0.15, ani.width=1300, ani.height=700,
movie.name="2003-murray-humanaccomplishment-region-proportions-bootstrap.gif", clean=FALSE)

The bootstrap animation suggests to me that while the very earliest time-periods are opaque and the Dark Ages difference between Europe & Asia may be somewhat higher or lower, the overall shape doesn’t change much.

efit <- auto.arima(subset(hdcount, select=c("Fl.decade", "Count"), Group=="European")$Count) afit <- auto.arima(subset(hdcount, select=c("Fl.decade", "Count"), Group=="Asian")$Count)
ofit <- auto.arima(subset(hdcount, select=c("Fl.decade", "Count"), Group=="Other")$Count) par(mfrow=c(3,1)) plot(forecast(efit), ylim=c(0,200)); axis(side=1, labels=decades, at=seq(1, length(decades))) plot(forecast(afit), ylim=c(0,200)); axis(side=1, labels=decades, at=seq(1, length(decades))) plot(forecast(ofit), ylim=c(0,200)); axis(side=1, labels=decades, at=seq(1, length(decades))) # https://imgur.com/a/qPL6a Program for non-spaced-repetition review of past written materials for serendipity & rediscovery: Archive Revisiter One reason to take notes/clippings and leave comments in stimulating discussions is to later benefit by having references & citations at hand, and gradually build up an idea from disparate threads and make new connections between them. For this purpose, I make extensive excerpts from web pages & documents I read into my Evernote clippings (functioning as a commonplace book), and I comment constantly on Reddit/LessWrong/#lesswrong etc. While expensive in time & effort, I often go back, months or years later, and search for a particular thing and expand & integrate it into another writing or expand it out to an entire essay of its own. (I also value highly not being in the situation where I believe something but I do not know why I believe it other than the conviction I read it somewhere, once.) This sort of personal information management using simple personal information managers like Evernote works well enough when I have a clear memory of what the citation/factoid was, perhaps because it was so memorable, or when the citations or comments are in a nice cluster (perhaps because there was a key phrase in them or I kept going back & expanding a comment), but it loses out on key benefits to this procedure: serendipity and perspective. As time passes, one may realize the importance of an odd tidbit or have utterly forgotten something or events considerably changed its meaning; in this case, you would benefit from revisiting & rereading that old bit & experiencing an aha! moment, but you don’t realize it. So one thing you could do is reread all your old clippings & comments, appraising them for reuse. But how often? And it’s a pain to do so. And how do you keep track of which you’ve already read? One thing I do for my emails is semi-annually I (try to) read through my previous 6 months of email to see what might need to be followed up on15 or mined for inclusion in an article. (For example, an ignored request for data, or a discussion of darknet markets with a journalist I could excerpt into one of my DNM articles so I can point future journalists at that instead.) This is already difficult, and it would be even harder to expand. I have read through my LessWrong comment history… once. Years ago. It would be more difficult now. Simply re-reading periodically in big blocks may work but is suboptimal: there is no interface easily set up to reread them in small chunks over time, no constraints which avoid far too many reads, nor is there any way to remove individual items which you are certain need never be reviewed again. Reviewing is useful but can be an indefinite timesink. (My sent emails are not too hard to review in 6-month chunks, but my IRC logs are bad - 7,182,361 words in #lesswrong alone - and my >38k Evernote clippings are worse; any lifestreaming will exacerbate the problem by orders of magnitude.) This is probably one reason that people who keep journals or diaries don’t reread Nor can it be crowdsourced or done by simply ranking comments by public upvotes (in the case of Reddit/LW/HN comments), because the most popular comments are ones you likely remember well & have already used up, and the oddities & serendipities you are hoping for are likely unrecognizable to outsiders. This suggests some sort of reviewing framework where one systematically reviews old items (sent emails, comments, IRC logs by oneself), putting in a constant amount of time regularly and using some sort of ever expanding interval between re-reads as an item becomes exhausted & ever more likely to not be helpful. Deconstructing Deathism - Answering Objections to Immortality, Mike Perry 2013 (note: this is an entirely different kind of problem than those considered in Freeman Dyson’s immortal intelligences in Infinite in All Directions, which are more fundamental), discusses something like what I have in mind in terms of an immortal agent trying to review its memories & maintain a sense of continuity, pointing out that if time is allocated correctly, it will not consume 100% of the agent’s time but can be set to consume some bounded fraction: It seems reasonable that past versions of the self would survive as we remember the events of times past, that is to say, our episodic memories, and this would have importance in our continuing to persist as what could be considered the same albeit also a changing, developing person. But in addition to this mnemonic reinforcement I imagine there would be a more general feeling of being a particular individual, an ambiance derived from but not referring to any specific past experiences. Ambiance alone would not be sufficient, I think, to make us who we are; episodic memories would also be necessary, yet it could considerably lessen the need for frequent recall and thus alleviate the problem of dilution. Another interesting thought is that certain items might consistently be consulted more frequently than others. (Indeed, would this not be expected?) In this way it would actually be possible to bypass the dilution effect and instead allow a fixed fraction of time for perusal of any given item, even as more items were added indefinitely. A simple way of doing this could be first to allow some fixed fraction of the time for day-to-day affairs and other non-archival work (prime time), and spend the rest of the time on perusal of personal archives (archive time). The exact apportioning of prime versus archive time is not important here, but it will be instructive to consider how the archive time itself might be subdivided. A simple, if overly simplistic, strategy would be to have half this time devoted to the first century’s records, half the remainder to the second century, and so on. (Since there would only be a finite number of centuries, there would be some unused archive time at the end, which could be spent as desired. Note, however, that in the limit of infinite total time covering infinitely many centuries, the usage of archive time would approach but not exceed 100%.) In this way, then, there would be a fixed fraction of archive time, $2-n$, spent on the _n_th century’s records, regardless of how many centuries beyond the nth were lived or how many records accumulated. True, this way of apportioning time might not be much good beyond a few centuries; only about one trillionth the total time would be spent on the 40th century, for instance, around $\frac{1}{300}$ sec per 100 years. (Possibly a lot could be covered even in this brief interval of about 3 million nanoseconds, however.) But the apportionment scheme could be adjusted. A more interesting and plausible, if slightly harder-to-describe scheme would be to choose a constant $c>0$ and allow the fraction $c \cdot (\frac{1}{n+c-1} - \frac{1}{n+c})$ to the _n_th-century records. It is easy to show that the time for all centuries will add up to 100% as before, whatever positive value of c we start with. Starting with $c=10$ will get 10% of the total time spent on the first century, with subsequent centuries receiving a diminishing share as before, but the rate of falloff will be much slower, so that the 40th century will still receive 0.4%, or about 5 months per 100 years, that is to say, 240 million nanoseconds per minute. If we suppose that our immortal settles eventually into a routine in which 10% of the time overall is archive time, there would be 24 million nanoseconds available each minute of life for the 40th century’s memories alone, if desired, with many other centuries getting more or less comparable or greater amounts of attention, and none omitted entirely. This, I think, makes at least a plausible case that a reasonable sense of one’s personal identity could be sustained indefinitely. In the above examples the greatest proportion of archive time falls to the earlier records, which might be fitting since these should be the most important as formative years for the prospective immortal, thus the most important for identity maintenance. (Memory recall would also naturally occur during prime time; the emphasis here could be on recent events, to maintain a balance overall.) In summary, then, we have considered ways that the problem of dilution might be successfully managed. Relatively infrequent perusal of memories might still suffice to maintain the necessary continuity with past versions of the self, or proper scheduling could stabilize the frequency of recall and bypass the dilution effect, or both. We see in any case that the problem is not what it may seem at first sight. We have no guarantee, of course, that it would not get out of bounds, but certainly some grounds for hope. So you could imagine some sort of software along the lines of spaced repetition systems like Anki/Mnemosyne/Supermemo which you spend, say, 10 minutes a day at, simply rereading a selection of old emails you sent, lines from IRC with n lines of surrounding context, Reddit & LW comments etc; with an appropriate backoff & time-curve, you would reread each item maybe 3 times in your lifetime (eg first after a delay of a month, then a year or two, then decades). Each item could come with a rating function where the user rates it as an important or odd-seeming or incomplete item and to be exposed again in a few years, or as totally irrelevant and not to be shown again - as for many bits of idle chit-chat, mundane emails, or intemperate comments is not an instant too soon! (More positively, anything already incorporated into an essay or otherwise reused likely doesn’t need to be resurfaced.) This wouldn’t be the same as a spaced repetition system which is designed to recall an item as many times as necessary, at the brink of forgetting, to ensure you memorize it; in this case, the forgetting curve & memorization are irrelevant and indeed, the priority here is to try to eliminate as many irrelevant or useless items as possible from showing up again so that the review doesn’t waste time. More specifically, you could imagine an interface somewhat like Mutt which reads in a list of email files (my local POP email archives downloaded from Gmail with getmail4, filename IDs), chunks of IRC dialogue (a grep of my IRC logs producing lines written by me +- 10 lines for context, hashes for ID), LW/Reddit comments downloaded by either scraping or API via the BigQuery copy up to 2015, and stores IDs, review dates, and scores in a database. One would use it much like a SRS system, reading individual items for 10 or 20 minutes, and rating them, say, upvote (this could be useful someday, show me this ahead of schedule in the future) / downvote (push this far off into the future) / delete (never show again). Items would appear on an expanding schedule. For example if one wanted to review items 4 times over the next 50 years (roughly my life expectancy), a schedule might be: round({t=0:4; t^6.981}) # [1] 0 1 126 2142 15958 So in 1 day, then a third of a year, then after 5.8 years, then after 43 years. Alternately, a geometric series might be a bit kinder and not too front-loaded: review <- function(n, r, a) { a * (1 - r^n) / (1 - r) } reviews <- function(n, r, a) { sapply(1:n, function(nn) { review(nn, r, a) }) } findR <- function (firstReview=31, n_total=3, years=50) { optimize(interval=c(0, 1000), f = function(r) { abs(sum(sapply(1:n_total, function(n){review(n, a=firstReview, r=r)})) - (365*years)) })$minimum }
findR(firstReview=30, n_total=4, years=50)
# [1] 7.728823216
round(reviews(4, 7.728823216, 30))
# [1]    30   262  2054 15904

The geometric series allows for easy incorporation of rating modifications: a downvote penalty might multiply r by 1.5, vs 0.5 for upvotes. This would also allow some input from statistical algorithms which predict upvote/downvote/delete and advances/delays items based on that, which would hopefully quickly learn to avoid idle chit-chat and short performative utterances and start to prioritize more interesting & unusual items. (For example, a good start might be a SVM on a bag-of-words version of each item’s text, and then as the dataset ratings expand, more complicated algorithms could be plugged in.)

As far as I know, some to-do/self-help systems have something like a periodic review of past stuff, and as I mentioned, spaced repetition systems do something somewhat similar to this idea of exponential revisits, but there’s nothing like this at the moment.

1. Richard Hamming, You and Your Research:

The three outstanding problems in physics, in a certain sense, were never worked on while I was at Bell Labs. By important I mean guaranteed a Nobel Prize and any sum of money you want to mention. We didn’t work on (1) time travel, (2) teleportation, and (3) antigravity. They are not important problems because we do not have an attack. It’s not the consequence that makes a problem important, it is that you have a reasonable attack. That is what makes a problem important.

2. The consensus among NICE’s economic advisers is that NICE should, generally, accept as cost effective those interventions with an incremental cost-effectiveness ratio of less than £20,000 per QALY and that there should be increasingly strong reasons for accepting as cost effective interventions with an incremental cost-effectiveness ratio of over £30,000 per QALY.

3. milk from herd that has been attested free from bovine tuberculosis. Tuberculosis infection from milk was a serious problem in the UK in this period; as J.B.S. Haldane remarks in his discussion of Vitamins (Possible Worlds And Other Essays 1927), I would sooner have my child run the risk of rickets or infantile scurvy from over-boiled milk than of tuberculosis from drinking it raw. I refer here to British milk—American is less tuberculous. Later elaborating in the The Fight With Tuberculosis chapter: Tuberculosis does not stand first on the list of causes of death in England, but it is the most serious, because it kills in infancy and prime of life…one quarter of the deaths of French children in their first two years are due to tubercle…he greatest single channel of infection is milk from tuberculous cows drunk in infancy or early childhood. But the vast majority even of well-to-do parents do not take the trouble to obtain Grade A or Grade A certified milk for their younger children. In many places it is, of course, not available, but it would be if an economic demand for it existed. And with no public opinion behind it in the matter the Government cannot be expected to legislate drastically in favour of pure milk. If science has not discovered a cure or an infallible preventive for tuberculosis, it has at least shown how the mortality could be greatly lowered. For the price of a cigar or a cinema a week you can protect your child against its most dangerous enemy. Is it worth while?

4. Note how much better identical twins are than siblings; sibling designs are valuable when we are hypothesizing about something in the shared-environment, and can be used for many purposes like showing that GWAS hits are not confounded by population stratification or that the harm from things like maternal smoking is overestimated by analyses ignoring genetic confounding, but since siblings are not all that similar, the gain is not particularly large. For some discussion of this topic from epidemiological perspectives, see Why are children in the same family so different from one another? & Smith 2011.

5. This renders the population estimate a bit off, but I couldn’t find any sources on the breakdown of Sephardic vs Ashkenazi in the USA in 1970 other than a comment that the latter were a vast majority. Since the Jewish Population Study was probably an undercount in not including all the people of Jewish descent, I’m hopeful those two biases cancel out.

6. The high IQ sample in Barbe 1964 would have been >8% Jewish, but the paper only reports the overall Jewishness, without specifying whether it’s 4% vs 12% or something like that.

7. Hollingworth & Rust 1937: The data of the present study were obtained early in 1933, the subjects being 36 boys and 19 girls, of the average age of 18 years 6 months. The IQ’s (S-B) of all had been taken in early childhood (9). The group ranged from 135-190 IQ (S-B), with a median at about 153 IQ (S-B). All but four of these young persons were Jewish, a factor which must be considered as of possible consequence (8, 14)…

8. Subotnik et al 1993, pg3-4:

The mean IQ of the Hunter sample was 157, or approximately 3.5 standard deviations above the mean, with a range of 122 to 196 on the L-M form. [Stanford-Binet Intelligence Scale, Form L-M (SBL-M)]

…Each class at Hunter College Elementary School from the years 1948 to 1960 contained about 50 students, yielding a total possible population of 600 graduates…35% of the total population of 1948-1960 HCES students (n=210) completed and returned study questionnaires

Religious Affiliation: The Hunter group is approximately 62% Jewish, although they describe themselves as Jews more in terms of ethnic identity than religious practice. The group, as a whole, is not religious.

Educational Attainments: Over 80% of the study participants held at least a Master’s degree. Furthermore, 40% of the women and 68% of the men held either a Ph.D, LL.B., J.D., or M.D. degree. Occupation and Income: Only two of the HCES women identified themselves primarily as homemakers. 53% were professionals, working as a teacher at the college or pre-college level, writer (journalist, author, editor), or psychologist. The same proportion of HCES men were professionals, serving as lawyers, medical doctors, or college teachers. The median income for men in 1988 was $75,000 (range =$500,000) and for women $40,000 (range =$169,000). Income levels were significantly different for men and women, even when matched by profession. For example, the median income for male college teachers or psychologists was $50,000 and for females,$30,000

9. My first attempt at it in JAGS went like this:

model_string <- '
model {
cutoffIQ <- 100 + 3.719016485*15

mu_asian ~ dnorm(105, 4^-2)
X_asian ~ dnorm(mu_asian, 15^-2) # T(cutoffIQ,)
X_frac_asian <- X_asian > cutoffIQ
P_asian <- 0.07 * (X_frac_asian / length(X_asian))
Y_asian ~ dbinom(P_asian, total)

# mu_white ~ dnorm(100, 4^-2)
# X_white ~ dnorm(mu_white, 15^-2) # T(cutoffIQ,)
# X_frac_white <- X_white > cutoffIQ
# P_white <- (1-0.07) * (X_frac_white / length(X_white))
# Y_white ~ dbinom(P_white, total)
}
'
library(runjags)
Y_asian=126
Y_white=418
total=579
model <- run.jags(model_string, data = list(Y_asian=Y_asian, Y_white=Y_white, total=total),
monitor=c("mu_asian", "mu_white"),
n.chains = getOption("mc.cores"), method="rjparallel")
summary(model)

But then I realized that X_frac_asian <- X_asian > cutoffIQ didn’t do what I thought it did and I needed to somehow draw a large number of samples, just like in the ABC simulation, and compare to the number after the truncation… or something.

10. Several passages mention that the students were repeatedly tested throughout their education, so it should’ve been entirely possible to look at IQ scores longitudinally and note how much they declined since admission, although it also seems possible that this decline will be masked by the constant testing leading to test-specific training and loss of validity in measuring g.

11. For comparison, Hunter High has always used only an exam for admission, aside from the grandfathered Elementary students, and a 2010 NYT article on a small flareup of the controversy prompted by a black-Hispanic student’s speech, says In 1995, the entering seventh-grade class was 12% black and 6% Hispanic, according to state data. This past year, it was 3% black and 1% Hispanic; the balance was 47% Asian and 41% white, with the other 8% of students identifying themselves as multiracial. The public school system as a whole is 70% black and Hispanic.

12. Consider, for instance, Hunter College Elementary School, perhaps the most competitive publicly funded school in the city. (This year, there were 36 applicants for each slot.) Four-year-olds won’t even be considered for admission unless their scores begin in the upper range of the 98th percentile of the Stanford-Binet Intelligence Scales, which costs \$275 to take. But if they’re accepted and successfully complete third grade (few don’t), they’ll be offered admission to Hunter College High School. And since 2002, at least 25% of Hunter’s graduating classes have been admitted to Ivy League schools. (In 2006 and 2007, that number climbed as high as 40.) Or take, as another example, Trinity School. In 2008, 36% of its graduates went to Ivy League schools. More than a third of those classes started there in kindergarten. 30% of Dalton’s graduates went to Ivies between 2005 and 2009, as did 39% of Collegiate’s and 34% of Horace Mann’s. Many of these lucky graduates wouldn’t have been able to go to these Ivy League feeders to begin with, if they hadn’t aced an exam just before kindergarten. And of course these advantages reverberate into the world beyond.

…Those who are bullish on intelligence tests argue they’re pure gauges of a child’s mental agility-immune to shifts in circumstance, immutable over the course of a lifetime. Yet everything we know about this subject suggests that there are considerable fluctuations in children’s IQs. In 1989, the psychologist Lloyd Humphreys, a pioneer in the field of psychometrics, came out with an analysis based on a longitudinal twin study in Louisville, Kentucky, whose subjects were regularly IQ-tested between ages 4 and 15. By the end of those eleven years, the average change in their IQs was ten points. That’s a spread with significant educational consequences. A 4-year-old with an IQ of 85 would likely qualify for remedial education. But that same child would no longer require it if, later on, his IQ shoots up to 95. A 4-year-old with an IQ of 125 would fall below the 130 cutoff for the G&T programs in most cities. Yet if, at some point after that, she scores a 135, it will have been too late. She’ll already have missed the benefit of an enhanced curriculum.

These fluctuations aren’t as odd as they seem. IQ tests are graded on a bell curve, with the average always being 100. (Definitions vary, but essentially, people with IQs of 110 to 120 are considered smart; 120 to 130, very smart; 130 is the favorite cutoff for gifted programs; and 140 starts to earn people the label of genius.) If a child’s IQ goes down, it doesn’t mean he or she has stopped making intellectual progress. It simply means that this child has made slower progress than some of his or her peers; the child’s relative standing has gone down. As one might imagine, kids go through cognitive spurts, just as they go through growth spurts. One of the classic investigations into the stability of childhood IQ, a 1973 study by the University of Pittsburgh’s Robert McCall and UC-San Diego’s Mark Appelbaum and colleagues, looked at 80 children who’d taken IQ tests roughly once a year between the ages of 2½ and 18. It showed that children’s intellectual trajectories were marked by slow increases or decreases, with inflection points around the ages of 6, 10, and 14, during which scores more sharply turned up or down. And when were IQs the least stable? Before the age of 6. Yet in New York we track most kids based on test scores they got at 4. (And we may not even be the worst offenders: As Po Bronson and Ashley Merryman note in their new book, NurtureShock, there are cities with preschools that require IQ tests off 2-year-olds.) How can you lock children into a specialized educational experience at so young an age? asks McCall. As soon as you start denying kids early, you penalize them almost progressively. Education and mental achievement builds on itself. It’s cumulative.

…Most researchers in the field of childhood development agree that the minds of nursery-school children are far too raw to be judged. Sally Shaywitz, author of Overcoming Dyslexia, is in the midst of a decades-long study that examines reading development in children. She says she couldn’t even use the reading data she’d collected from first-graders for some of the longitudinal analyses. It simply wasn’t stable, she says. I tell her that most New York City schools don’t share this view. A young brain is a moving target, she replies. It should not be treated as if it were fixed.

…In Genius Revisited, Rena Subotnik, director of the American Psychological Association’s Center for Gifted Education Policy, undertook a similar study, with colleagues, looking at Hunter elementary-school alumni all grown up. Their mean IQs were 157. They were lovely people, she says, and they were generally happy, productive, and satisfied with their lives. But there really wasn’t any wow factor in terms of stellar achievement.

…If you’re looking for practical answers though, Plucker, of Indiana, has a modest proposal. He suggests that schools assess children at an age when IQs get more stable. And in fact, that’s just what City and Country, one of Manhattan’s more progressive schools, does. Standardized tests aren’t required of their applicants until they’re 7 or older. That way, the kids are further along in their schooling, explains Elise Clark, the school’s admissions director. They’re used to an academic setting, they can handle a test-taking situation, and overall, we consider the results more reliable.

13. Subotnik in particular seems to have not expected such regression to the mean (Subotnik et al 2011):

In 2003, Subotnik commented on the surprise she had felt a decade before at realizing that graduates of an elite program for high-IQ children had not made unique contributions to society beyond what might be expected from their family SES and the high-quality education they received (see Subotnik, Kassan, et al., 1993), and posed the following question to readers: Can gifted children grown up claim to be gifted adults without displaying markers of distinction associated with their abilities? (Subotnik, 2003, p. 14).

…However, the disconnect between childhood giftedness and adult eminence (Cross & Coleman, 2005; Dai, 2010; Davidson, 2009; Freeman, 2010; Subotnik et al. Hollinger & Fleming, 1992; Simonton, 1991, 1998; Subotnik & Rickoff, 2010; VanTassel-Baska, 1989), as well as the outcomes of individuals who receive unexpected opportunities (Gladwell, 2008; Syed, 2010), suggest that there is a much larger base of talent than is currently being tapped.

14. Strictly speaking we could probably do a parametric bootstrap by counting per decade, and treating each one as a Poisson estimation, and generating random deviates, but that runs into issues with decades with 0 figures (the MLE might estimate a rate of 0, but we know that’s wrong).

15. I’ve been experimenting with Boomerang to reduce the problem of non-followups by setting ping me if no reply within 1 month alerts on my sent emails.