Archive for the ‘Surgeon Scorecard’ Category

Summing up my criticism against Propublica’s Scorecard

August 26, 2015

The bottom-line: The major justification for my substantiated (or at least so I think) rant against the use of shrinkage in this case is the elephant in the room that no-one wants to talk about: the substantially limited information (avg number of complications per individual < 50) for otherwise infrequent events (avg complication rate <5%). In such a situation, shrinkage will make everyone look like everyone else, limiting the ability to draw meaningful complications by looking at the values of the random effects. In fact, PP had to rely on post hock classifications of the good, the bad and the ugly to overcome the unavoidable shrinkage towards the mean and overcome the lack of information that could distinguish one surgeon from another. Similar points (with more examples from the actual score card) were made by Ewen Harrison.

(You may stop reading now – or you may read past the rant in the next paragraph)

Rant Mode On: Another criticism that I received, is that I fail to understand random effects modeling, which I personally thought it was funny because one of my recent papers actually says one should scrap the Cox survival model for generalized additive models (which are just generalized linear mixed models in disguise).  In any case, assuming that my understanding of mixed models is poor, maybe my conclusion that these models are to be preferred for such applications is also problematic?

Since one may consider my vote-of-confidence-to-the-almighty-mix-model v.s. my criticism of their application in the Scorecard project, subtle evidence for paranoid schizophrenia let me sum up how one can simultaneously hold these beliefs:

Even though mixed models are to be preferred especially as more data accumulate(a point I make clear in the 3 blog posts I wrote), no modeling could overcome the severe lack of events in the scorecard database.

There are other technical issues that one could “rant” about e.g. how were the hospital and surgeon effects were combined, but I will not digress any further.

Rant mode off

Back to criticism: In my opinion a major reason that mixed models were used is the potentially large number of surgeons with zero complications in their limited observation records*. The use of shrinkage models allow one to report performance for such surgeons, and generate nice colorful graphs (with one and possibly two decimal points) about their performance instead of reporting a NaN (Not A Number). Incidentally, the shrinkage of the individual surgeon effects all the way towards the population mean is the mixed model’s way of telling you the same thing: since it could not estimate how these people actually performed, it might as well give back the only answer consistent with the model applied, i.e. everybody is (almost) like everybody else and thus everyone is average.

*Curiously this information, i.e. the number of surgeons without any complication in the source database is not provided, although I’d have thought it would be important to report this number in the shake of transparency.

Word of caution: For surgeons without any complications in the database, the actual information comes not from their complications but from the number of uncomplicated surgeries they performed. This raises the question of inclusion effects (a point Andrew Gelman thoroughly address in his book) and the associated selection biases inherent in comparing surgeons who accept v.s. those who do not accept Medicare (or accept too little of it) and the corresponding social determinants of health outcomes (beautifully explained here).

A non-statistical criticism: As a nephrologist, I find it amusingly insulting that Acute Kidney Injury (coded as Acute Renal Failure) was not considered a surgical complication, even though it certainly complicates surgeries. Surgery is in fact the leading cause of “hospital-acquired” AKI (40% of cases are preceded by a surgical procedure). But maybe, as Benjamin Davies pointed out, the real end-points are not really measured by the Scorecard. Or I could be just delusional when I point out to surgical colleagues that the selection of analgesia, fluid management and pre as well as post perioperative care DO play a role for some complications. Or it could be just the anesthesiologists’s fault 🙂


The little mixed model that could, but shouldn’t be used to score surgical performance

July 30, 2015

The Surgeon Scorecard

Two weeks ago, the world of medical journalism was rocked by the public release of ProPublica’s Surgeon Scorecard. In this project ProPublica “calculated death and complication rates for surgeons performing one of eight elective procedures in Medicare, carefully adjusting for differences in patient health, age and hospital quality.”  By making the dataset available through a user friendly portal, the intended use of this public resource was to “use this database to know more about a surgeon before your operation“.

The Scorecard was met with great enthusiasm and coverage by non-medical media. headline “nutshelled” the Scorecard as a resource that “aims to help you find doctors with lowest complication rates“. A (?tongue in cheek) NBC headline tells us the scorecard “It’s complicated“. On the other hand the project was not well received by my medical colleagues. John Mandrola gave it a failing grade in Medscape. Writing at, Jeffrey Parks called it a journalistic low point for ProPublica. Jha Shaurabh pointed out a potential paradox  in a statistically informed, easy to read and highly entertaining piece. In this paradox, the surgeon with the higher complication case who takes high risk patients from a disadvantaged socio-economic environment, may actually be the surgeon one wants to perform one’s surgery! Ed Schloss summarized the criticism (in the blogosphere and twitter) in an open letter and asked for peer review of the Scorecard methodology.

The criticism to date has largely focused on the potential for selection effects (as the Scorecard is based on Medicare data, and does not include data from private insurers), the incomplete adjustment for confounders, the paucity of data for individual surgeons, the counting of complications and re-admission rates, decisions about risk category classification boundaries and even data errors (ProPublica’s response arguing that the Scorecard matters may be found here). With a few exceptions (e.g. see Schloss’s blogpost in which the complexity of the statistical approach is mentioned) the criticism of the statistical approach (including my own comments in twitter) has largely focused on these issues.

On the other hand, the underlying statistical methodology (here and there) that powers the Scorecard has not received much attention. Therefore I undertook a series of simulation experiments to explore the impact of the statistical procedures on the inferences afforded by the Scorecard.

The mixed model that could – a short non-technical summary of ProPublica’s approah

ProPublica’s approach to the scorecard is based on logistic regression model, in which individual surgeon (and hospital) performance (probability of suffering a complication) is modelled using Gaussian random effects, while patient level characteristics that may act as confounders are adjusted for, using fixed effects. In a nutshell this approach implies fitting a model of the average complication rate that is function of the fixed effects (e.g. patient age) for the entire universe of surgeries  performed in the USA. Individual surgeon and hospital factors modify this complication rate, so that a given surgeon and hospital will have an individual  rate that varies around the population average. These individual surgeon and hospital factors are constrained to follow Gaussian, bell-shaped distribution when analyzing complication data. After model fitting, these predicted random effects are used to quantify and compare surgical performance. A feature of mixed modeling approaches is the unavoidable shrinkage of the raw complication rate towards the population mean. Shrinkage implies that the dynamic range of the  actually observed complication rates is compressed. This is readily appreciated in the figures generated by the ProPublica analytical team:


In their methodological white paper the ProPublica team notes:

While raw rates ranged from 0.0 to 29%, the Adjusted Complication Rate goes from 1.1 to 5.7%. …. shrinkage is largely a result of modeling in the first place, not due to adjusting for case mix. This shrinkage is another piece of the measured approach we are taking: we are taking care not to unfairly characterize surgeons and hospitals.”

These features should alert us that something is going on. For if a model can distort the data to such a large extent, then the model should be closely scrutinized before being accepted. In fact, given these observations,  it is possible that one mistakes the noise from the model for the information hidden in the empirical data. Or, even more likely, that one is not using the model in the most productive manner.

Note that these comments  should not be interpreted as a criticism against the use of mixed models in general, or even for the particular aspects of the Scorecard project. They are rather a call for re-examining the modeling assumptions and for gaining a better understanding of the model “mechanics of prediction” before releasing the Kraken to the world.

The little mixed model that shouldn’t

There are many technical aspects one could potentially misfire in a Generalized Linear Mixed Model for complication rates. Getting the wrong shape of the random effects distribution is of may or may not be of concern (e.g. assuming it is bell shaped when it is not). Getting the underlying model wrong, e.g. assuming the binomial model for complication rates while a model with many more zeros (a zero inflated model) may be more appropriate, is yet another potential problem area. However, even if  these factors are not operational, then one may still be misled when using the results of the model. In particular, the major area of concern for such models is the cluster size: the number of observations per individual random effect (e.g. surgeon) in the dataset. It is this factor, rather than the actual size of the dataset that determines the precision in the individual random affects. Using a toy example, we show that the number of observations per surgeon typical of the Scorecard dataset, leads to predicted  random effects that may be far from their true value. This seems to stem from the non-linear nature of the logistic regression model. As we conclude in our first technical post:

  • Random Effect modeling of binomial outcomes require a substantial number of observations per individual (in the order of thousands) for the procedure to yield estimates of individual effects that are numerically indistinguishable  from the true values.


Contrast this conclusion to the cluster size in the actual scorecard:

Procedure Code N (procedures) N(surgeons) Procedures per surgeon
51.23 201,351 21,479 9.37
60.5 78,763 5,093 15.46
60.29 73,752 7,898 9.34
81.02 52,972 5,624 9.42
81.07 106,689 6,214 17.17
81.08 102,716 6,136 16.74
81.51 494,576 13,414 36.87
81.54 1,190,631 18,029 66.04
Total 2,301,450 83,887 27.44

In a follow up simulation study we demonstrate that this feature results in predicted individual effects that are non-uniformly shrank towards their average value. This compromises the ability of mixed model predictions to separate the good from the bad “apples”.

In the second technical post, we undertook a simulation study to understand the implications of over-shrinkage for the Scorecard project. These are best understood by a numerical example from one of simulated datasets. To understand this example one should note that the individual random effects have the interpretation of (log-) odds ratios. Hence, the difference in these random effects when exponentiated yield the odds ratio of suffering a complication in the hands of a good relative to a bad surgeon. By comparing these random effects for good and bad surgeons who are equally bad (or good) relative to the mean (symmetric quantiles around the median), one can get an idea of the impact of using the predicted random effects to carry out individual comparisons.

Good Bad Quantile (Good) Quantile (Bad) True OR Pred OR Shrinkage Factor
-0.050 0.050 48.0 52.0 0.905 0.959 1.06
-0.100 0.100 46.0 54.0 0.819 0.920 1.12
-0.150 0.150 44.0 56.0 0.741 0.883 1.19
-0.200 0.200 42.1 57.9 0.670 0.847 1.26
-0.250 0.250 40.1 59.9 0.607 0.813 1.34
-0.300 0.300 38.2 61.8 0.549 0.780 1.42
-0.350 0.350 36.3 63.7 0.497 0.749 1.51
-0.400 0.400 34.5 65.5 0.449 0.719 1.60
-0.450 0.450 32.6 67.4 0.407 0.690 1.70
-0.500 0.500 30.9 69.1 0.368 0.662 1.80
-0.550 0.550 29.1 70.9 0.333 0.635 1.91
-0.600 0.600 27.4 72.6 0.301 0.609 2.02
-0.650 0.650 25.8 74.2 0.273 0.583 2.14
-0.700 0.700 24.2 75.8 0.247 0.558 2.26
-0.750 0.750 22.7 77.3 0.223 0.534 2.39
-0.800 0.800 21.2 78.8 0.202 0.511 2.53
-0.850 0.850 19.8 80.2 0.183 0.489 2.68
-0.900 0.900 18.4 81.6 0.165 0.467 2.83
-0.950 0.950 17.1 82.9 0.150 0.447 2.99
-1.000 1.000 15.9 84.1 0.135 0.427 3.15
-1.050 1.050 14.7 85.3 0.122 0.408 3.33
-1.100 1.100 13.6 86.4 0.111 0.390 3.52
-1.150 1.150 12.5 87.5 0.100 0.372 3.71
-1.200 1.200 11.5 88.5 0.091 0.356 3.92
-1.250 1.250 10.6 89.4 0.082 0.340 4.14
-1.300 1.300 9.7 90.3 0.074 0.325 4.37
-1.350 1.350 8.9 91.1 0.067 0.310 4.62
-1.400 1.400 8.1 91.9 0.061 0.297 4.88
-1.450 1.450 7.4 92.6 0.055 0.283 5.15
-1.500 1.500 6.7 93.3 0.050 0.271 5.44
-1.550 1.550 6.1 93.9 0.045 0.259 5.74
-1.600 1.600 5.5 94.5 0.041 0.247 6.07
-1.650 1.650 4.9 95.1 0.037 0.236 6.41
-1.700 1.700 4.5 95.5 0.033 0.226 6.77
-1.750 1.750 4.0 96.0 0.030 0.216 7.14
-1.800 1.800 3.6 96.4 0.027 0.206 7.55
-1.850 1.850 3.2 96.8 0.025 0.197 7.97
-1.900 1.900 2.9 97.1 0.022 0.188 8.42
-1.950 1.950 2.6 97.4 0.020 0.180 8.89
-2.000 2.000 2.3 97.7 0.018 0.172 9.39
-2.050 2.050 2.0 98.0 0.017 0.164 9.91

From this table it can be seen that predicted odds ratios are always larger than the true ones. The ratio of these odds ratios (the shrinkage factor) is larger, the more extreme comparisons are contemplated.

In summary, the use of the random effects models for the small cluster sizes (number of observations per surgeon) is likely to lead to estimates (or rather predictions) of individual effects that are smaller than their true values. Even though one should expect the differences to decrease with larger cluster sizes, this is unlikely to happen in real world datasets (how often does one come across a surgeon who has performed 1000s of operation of the same type before they retire?). Hence, the comparison of  surgeon performance based on these random effect predictions is likely to be misleading due to over-shrinkage.

Where to go from here?

ProPublica should be congratulated for taking up such an ambitious, and ultimately useful project. However, the limitations of the adopted approach should make one very skeptical about accepting the inferences from their modeling tool. In particular, the small number of observations per surgeon limits the utility of the predicted random effects to directly compare surgeons due to over-shrinkage. Further studies are required before one could use the results of mixed effects modeling for this application. Based on some limited simulation experiments (that we do not present here), it seems that relative rankings of surgeons may be robust measures of surgical performance, at least compared to the absolute rates used by the Scorecard. Adding my voice to that of Dr Schloss, I think it is time for an open and transparent dialogue (and possibly a “crowdsourced” research project) to better define the best measure of surgical performance given the limitations of the available data. Such a project could also explore other directions, e.g. the explicit handling of zero inflation and even go beyond the ubiquitous bell shaped curve. By making the R code available, I hope that someone (possibly ProPublica) who can access more powerful computational resources can perform more extensive simulations. These may better define other aspects of the modeling approach and suggest improvements in the scorecard methodology. In the meantime, it is probably a good idea not to exclusively rely on the numerical measures of the scorecard when picking up the surgeon who will perform your next surgery.



Empirical bias analysis of random effects predictions in linear and logistic mixed model regression

July 30, 2015

In the first technical post in this series, I conducted a numerical investigation of the biasedness of random effect predictions in generalized linear mixed models (GLMM), such as the ones used in the Surgeon Scorecard, I decided to undertake two explorations: firstly, the behavior of these estimates as more and more data are gathered for each individual surgeon and secondly whether the limiting behavior of these estimators critically depends on the underlying GLMM family. Note that the first question directly assesses whether the random effect estimators reflect the underlying (but unobserved) “true” value of the individual practitioner effect in logistic regression models for surgical complications. On the other hand the second simulation examines a separate issue, namely whether the non-linearity of the logistic regression model affects the convergence rate of the random effect predictions towards their true value.

For these simulations we will examine three different ranges of dataset sizes for each surgeon:

  • small data (complication data from between 20-100 cases/ surgeon are available)
  • large data (complications from between 200-1000 cases/surgeon)
  • extra large data (complications from between 1000-2000 cases/surgeon)

We simulated 200 surgeons (“random effects”) from a normal distribution with a mean of zero and a standard deviation of 0.26, while the population average complication rate was set t0 5%. These numbers were chosen to reflect the range of values (average and population standard deviation) of the random effects in the Score Card dataset, while the use of 200 random effects was a realistic compromise with the computational capabilities of the Asus Transformer T100 2 in 1 laptop/tablet that I used for these analyses.

The following code was used to simulate the logistic case for small data (the large and extra large cases were simulated by changing the values of the Nmin and Nmax variables).

## helper functions
logit<-function(x) log(x/(1-x))
invlogit<-function(x) exp(x)/(1+exp(x))

## simulate cases
simcase<-function(N,p) rbinom(N,1,p)
## simulation scenario
pall<-0.05; # global average
Nsurgeon<-200; # number of surgeons
Nmin<-20; # min number of surgeries per surgeon
Nmax<-100; # max number of surgeries per surgeon

## simulate individual surgical performance
## how many simulations of each scenario
set.seed(123465); # reproducibility
ind<-rnorm(Nsurgeon,0,.26) ; # surgical random effects
logitind<-logit(pall)+ind ; # convert to logits
pind<-invlogit(logitind); # convert to probabilities
Nsim<-sample(Nmin:Nmax,Nsurgeon,replace=TRUE); # number of cases per surgeon

complications<-data.frame(,mapply(simcase,Nsim,pind,SIMPLIFY=TRUE)),,mapply(function(i,N) rep(i,N),1:Nsurgeon,Nsim)))


A random effect and fixed effect model were fit to these data (the fixed effect model is simply a series of independent fits to the data for each random effect):

## Random Effects


## Fixed Effects

for(i in 1:Nsurgeon) {

The corresponding Gaussian GLMM cases were simulated by making minor changes to these codes. These are shown below:

simcase<-function(N,p) rnorm(N,p,1)



The predicted random effects were assessed against the simulated truth by smoothing regression splines. In these regressions, the intercept yields the bias of the average of the predicted random effects vis-a-vis the truth, while the slope of the regression quantifies the amount of shrinkage effected by the mixed model formulation. For unbiased estimation not only would we like the intercept to be zero, but also the slope to be equal to one. In this case, the predicted random effect would be equal to its true (simulated) value. Excessive shrinkage would result in a slope that is substantially different from one. Assuming that the bias (intercept) is not different from zero, the relaxation of the slope towards one quantifies the consistency and the bias (or rather its rate of convergence) of these estimators using simulation techniques (or so it seems to me).

The use of smoothing (flexible), rather than simple linear regression, to quantify these relationships does away with a restrictive assumption: that the amount of shrinkage is the same throughout the range of the random effects:

## smoothing spline (flexible) fit
## linear regression

The following figure shows the results of the flexible regression (black with 95% CI, dashed black) v.s. the linear regression (red) and the expected (blue) line (intercept of zero, slope of one).

Predicted v.s. simulated random effects for logistic and linear mixed regression as a function of the number of observations per random effect (cluster size)

Predicted v.s. simulated random effects for logistic and linear mixed regression as a function of the number of observations per random effect (cluster size)

Several observations are worth noting in this figure.
, the flexible regression was indistinguishable from a linear regression in all cases; hence the red and black lines overlap. Stated in other terms, the amount of shrinkage was the same across the range of the random effect values.
Second, the intercept in all flexible models was (within machine precision) equal to zero. Consequently, when estimating a group of random effects their overall mean is (unbiasedly) estimated.
Third, the amount of shrinkage of individual random effects appears to be excessive for small sample sizes (i.e. few cases per surgeon). Increasing the number of cases decreases the shrinkage, i.e. the black and red lines come closer to the blue line as N is increased from 20-100 to 1000-2000. Conversely, for small cluster sizes the amount of shrinkage is so excessive that one may lose the ability to distinguish between individuals with very different complication rates. This is reflected by a regression line between the predicted and the simulated random effect value that is nearly horizontal.
Fourth, the rate of convergence of the predicted random effects to their true value critically depends upon the linearity of the regression model. In particular, the shrinkage of logistic regression model with 1000-2000 observations per case is almost the same at that of a linear model with 20-100 for the parameter values considered in this simulation.

An interesting question is whether these observations (overshrinkage of random effects from small sample sizes in logistic mixed regression) reflects the use of random effects in modeling, or whether they are simply due to the interplay between sample size and the non-linearity of the statistical model. Hence, I turned to fixed effects modeling of the same datasets. The results of these analyses are summarized in the following figure:

Difference between fixed effect estimates of random effects(black histograms) v.s. random effects predictions (density estimators: red lines) relative to their simulated (true) values

Difference between fixed effect estimates of random effects(black histograms) v.s. random effects predictions (density estimators: red lines) relative to their simulated (true) values

One notes that the distribution of the differences between the random and fixed effects relative to the true (simulated) values is nearly identical for the linear case (second row). In other words, the use of the implicit constraint of the mixed model, offers no additional advantage when estimating individual performance in this model. On the other hand there is a value in applying mixed modeling techniques for the logistic regression case. In particular, outliers (such as those arising for small samples) are eliminated by the use of random effect modeling. The difference between the fixed and the random effect approach progressively decreases for large sample sizes, implying that the benefit of the latter approach is lost for “extra large” cluster sizes.

One way to put these differences into perspective is to realize that the random effects for the logistic model correspond to log-odd ratios, relative to the population mean. Hence the difference between the predicted random effect and its true value, when exponentiated, corresponds to an Odd Ratio (OR). A summary of the odds ratios over the population of the random effects as a function of cluster size is shown below.

Metric 20-100  200-1000 1000-2000
Min    0.5082   0.6665    0.7938
Q25    0.8901   0.9323    0.9536
Median 1.0330   1.0420    1.0190 
Mean   1.0530   1.0410    1.0300  
Q75    1.1740   1.1340    1.1000   
Max    1.8390   1.5910    1.3160 


Even though the average Odds Ratio is close to 1, a substantial number of predicted random effects are far from the true value and yield ORs that are greater than 11% in either direction for small cluster sizes. These observations have implications for the Score Card (or similar projects): if one were to use Random Effects modeling to focus on individuals, then unless the cluster sizes (observations per individual) are substantial, one would run a substantial risk of misclassifying individuals, even though one would be right on average!

One could wonder whether these differences between the simulated truth and the predicted random effects arise as a result of the numerical algorithms of the lme4 package. The latter was used by both the Surgeon Score Card project and our simulations so far and thus it would be important to verify that it performs up to specs. The major tuning variable for the algorithm is the order of the Adaptive Gaussian Quadrature (argument nAGQ). We did not find any substantial departures when the order of the quadrature was varied from 0 to 1 and 2. However, there is a possibility that the algorithm fails for all AGQ orders as it has to calculate probabilities that are numerically close to the boundary of the parameter space. We thus decided to fit the same model from a Bayesian perspective using Markov Chain Monte Carlo (MCMC) methods. The following code will fit the Bayesian model and graphs the true values of the effects used in the simulated dataset against the Bayesian estimates (the posterior mean) and also the lme4 predictions. The latter tend to be numerically close to the posterior mode of the random effects when a Bayesian perspective is adopted.

## Fit the mixed effects logistic model from R using openbugs

fitBUGS = glmmBUGS(ev ~ 1, data=complications, effects="id", family="bernoulli")
startingValues = fitBUGS$startingValues
fitBUGSResult = bugs(fitBUGS$ragged, getInits, = names(getInits()),
model.file="model.txt", n.chain=3, n.iter=6000, n.burnin=2000, n.thin=10,

fitBUGSParams = restoreParams(fitBUGSResult , fitBUGS$ragged)
sumBUGS<-summaryChain(fitBUGSParams )
checkChain(fitBUGSParams )

## extract random effects


## plot against the simulated (true) effects and the lme4 estimates


The following figure shows the histogram of the true values of the random effects (black), the frequentist(lme4) estimates (red) and the Bayesian posterior means (blue).



It can be appreciated that both the Bayesian estimates and the lme4 predictions demonstrate considerable shrinkage relative to the true values for small cluster sizes (20-100). Hence, an lme4 numerical quirk seems an unlikely explanation for the shrinkage observed in the simulation.

Summing up:

  • Random Effect modeling of binomial outcomes require a substantial number of observations per individual (cluster size) for the procedure to yield estimates of individual effects that are numerically indistinguishable  from the true values
  • Fixed effect modeling is even worse an approach for this problem
  • Bayesian fitting procedures do not appear to yield numerically different effects from their frequentist counterparts

These features should raise the barrier for accepting a random effects logistic modeling approach when the focus is on individual rather than population average effects. Even though the procedure is certainly preferable to fixed effects regression, the direct use of the value of the predicted individual random effects as an effect measure will be problematic for small cluster sizes (e.g. a small number of procedures per surgeon). In particular, a substantial proportion of these estimated effects is likely to be far from the truth even if the model is unbiased on the average. These observations are of direct relevance to the Surgical Score Card in which the number of observations per surgeon were far lower than the average value in our simulations: 60 (small), 600 (large) and 1500 (extra large):

Procedure Code N (procedures) N(surgeons) Procedures per surgeon
51.23 201,351 21,479 9.37
60.5 78,763 5,093 15.46
60.29 73,752 7,898 9.34
81.02 52,972 5,624 9.42
81.07 106,689 6,214 17.17
81.08 102,716 6,136 16.74
81.51 494,576 13,414 36.87
81.54 1,190,631 18,029 66.04
Total 2,301,450 83,887 27.44