Regression coefficients if 1 d.f. per factor, no interaction
Not standardized regression coefficients
Many programs print meaningless estimates such as effect of increasing age\(^2\) by one unit, holding age constant
Need to account for nonlinearity, interaction, and use meaningful ranges
For monotonic relationships, estimate \(X\hat{\beta}\) at quartiles of continuous variables, separately for various levels of interacting factors
Subtract estimates, anti-log, e.g., to get inter-quartile-range odds or hazards ratios. Base C.L. on s.e. of difference. See Figure 21.4.
Partial effect plot: Plot effect of each predictor on \(X\beta\) or some transformation. See Figure 21.2. See also Karvanen & Harrell (2009). Fox and Weisberg have other excellent approaches to displaying predictor effects
Forming contrasts through differences in predicted values is the most general and easy way to estimate effects of interest in a model. This is because
One needs only to specify two or more sets of predictor settings to contrast
Predictor coding, nonlinear terms, and interactions are automatically taken into account
Unspecified predictors can default to reference values such as medians for continuous predictors and modes for categorical ones. The values used for unspecified predictors only matter if they interact with predictors that are changing in the contrast.
Approximate methods are available for getting simultaneous confidence intervals over a whole sequence of contrasts (e.g., simultaneous confidence bands for an entire range of ages) when contrasts produce more than one estimate
The approach extends to any order of differences. E.g., interaction contrasts are formed with double differences.
Contrasts allow assessment of linearity over a given predictor range in a spline fit
Contrasts allow assessment of “partial interactions”, i.e., whether there is an interaction between one predictor and a subset of the levels of another predictor
The R predict function provides the \(X\) (design matrix) that goes along with each requested prediction, and simple matrix algebra provides needed covariance matrices once differences in design matrices are formed
On the latter point, a single difference contrast such as one comparing 30 year old males with 40 year old males would generate two design matrices \(X_{1}, X{2}\), and the contrast \(X_{1}\beta - X_{2}\beta = (X_{1} - X_{2})\beta\).
Consider a series of examples in which predicted values from a fitted model are symbolized by a function \(f\) of the predictor settings, with continuous predictors \(A\) and \(B\) and a categorical predictor \(C\) with levels \(i, j, k\). Let the reference (detault) values of the predictors be \(a\) (median of \(A\)), \(b\) (median of \(B\)), and \(c\) (modal (most frequent) category of \(C\)).
The term “average slope” used below will be the slope if the predictor has a linear effect, otherwise it is the average slope over the indicated interval. See here for more information.
Single Difference Contrasts
Average slope of \(A\) over \(A \in [1, 2]\): \(f(2, b, c) - f(1, b, c)\)
Difference between categories \(j\) and \(k\) in \(C\): \(f(a, b, j) - f(a, b, k)\)
Effect of an entire range of predictor values against a single reference value: \(f(1:10, b, c) - f(5, b, c)\) (a simultaneous confidence band would be useful here)
Double Difference Contrasts
Linearity in \(A\) over \(A \in [1, 3]\): \(f(3, b, c) - f(2, b, c) - [f(2, b, c) - f(1, b, c)] = f(3, b, c) - 2f(2, b, c) + f(1, b, c)\)
Interaction between \(A\) and \(C\) with respect to only two levels of \(C\): \(f(2, b, j) - f(1, b, j) - [f(2, b, k) - f(1, b, k)]\)
Interaction between \(A\) and \(B\) when \(A\) varies from 1 to 3 and \(B\) varies from 5 to 6: \(f(3, 5, c) - f(1, 5, c) - [f(3, 6, c) - f(1, 6, c)]\)
Effect of \(C=i\) vs. \(C=j\) over a regular grid of values of \(B\): \(f(a, 5:15, i) - f(a, 5:15, j)\) (with possible simultaneous band; this is often a better way of getting hazard ratios in a Cox proportional hazards model, as the user has control of the reference value)
Triple Difference Contrasts
Third-order interactions
Assessment of nonlinearity in the \(A\times B\) interaction (differences in double differences)
The R rms package contrast function (full name contrast.rms) provides predicted values, differences, and double difference contrasts. The latter is specified by providing four lists of predictor settings.
5.1.2 Indexes of Model Performance
Error Measures
Central tendency of prediction errors
Mean absolute prediction error: mean \(|Y - \hat{Y}|\)
calibration–in–the–large: average \(\hat{Y}\) vs. average \(Y\)
high-resolution calibration curve (calibration–in–the–small). See ?fig-titanic-calibrate.
calibration slope and intercept
maximum absolute calibration error
mean absolute calibration error
0.9 quantile of calibration error
CDE
See Van Calster et al. (2016) for a nice discussion of different levels of calibration stringency and their relationship to likelihood of errors in decision making.
\(g\)–Index
Based on Gini’s mean difference
mean over all possible \(i \neq j\) of \(|Z_{i} - Z_{j}|\)
interpretable, robust, highly efficient measure of variation
\(g =\) Gini’s mean difference of \(X_{i}\hat{\beta} = \hat{Y}\)
Example: \(Y=\) systolic blood pressure; \(g = 11\)mmHg is typical difference in \(\hat{Y}\)
Independent of censoring etc.
For models in which anti-log of difference in \(\hat{Y}\) represent meaningful ratios (odds ratios, hazard ratios, ratio of medians): \(g_{r} = \exp(g)\)
For models in which \(\hat{Y}\) can be turned into a probability estimate (e.g., logistic regression): \(g_{p} =\) Gini’s mean difference of \(\hat{P}\)
These \(g\)–indexes represent e.g. “typical” odds ratios, “typical” risk differences
Can define partial \(g\)
FG
5.1.3 Relative Explained Variation
In OLS linear models \(R^2\) is an excellent measure of explained variation in \(Y\)
Fraction of variance of \(Y\) that is explained by \(X\)
SSR = \(\sum_{i=1}^{n} (\hat{Y}_{i} - \bar{\hat{Y}})^2\) = regression sum of squares
proportional to variance of \(\hat{Y}\)
SSE = \(\sum_{i=1}^{n} (Y_{i} - \hat{Y}_{i})^2\) = error sum of squares = residual sum of squares
proportional to error variance
SST = \(\sum_{i=1}^{n} (Y_{i} - \bar{Y})^2\) = total sum of squares
proportion to variance of \(Y\)
SST = SSR + SSE
\(R^{2} =\) SSR/SST = 1 - SSE/SST
For each part of the model can compute its partial SSR = loss in SSR after refitting the model upon removing the part of the model being assessed
Partial \(R^{2}\) = (partial SSR)/SST
Relative partial \(R^{2} = \frac{\text{partial } R^{2}}{R^{2}} =\) partial SSR divided by full model SSR
Relative partial \(R^{2}\) = portion of explainable (by \(X\)) variation in \(Y\) that is explained by one or more terms in the model
Relative explained variation: REV
REVs do not need to sum to 1.0
collinearities
computing REVs for overlapping model terms
When a predictor interacts with another predictor, the predictor’s REV is taken from the combined predictive information in
the predictor main effect (including any nonlinear terms)
any interaction terms involving that effect
Extension to Other Models
For models fitted using maximum likelihood estimation, can compute the partial likelihood ratio \(\chi^2\) statistic for each set and divide that by the total model LR \(\chi^2\)
This is an excellent measure but requires more computation and doesn’t extend to models not having a likelihood or to Bayesian modeling
Need a general purpose REV measure that works on virtually all models using all estimation techniques
Letting \(\hat{Y} = X\hat{\beta}\) be the estimated linear predictor from the model
Use the same design matrix \(X\) used to get \(\hat{Y}\)
Predict \(\hat{Y}\) from \(X\) using OLS
The \(R^{2}\) from this model is 1.0
SSE = 0; ignore this and take variance-covariance matrix as \((X'X)^{-1}\) since SSR is used only on a relative basis
REV = \(\frac{\text{partial SSR }}{\text{SSR}}\)
Identical to relative \(R^2\) if original model was OLS
Easily lends itself to uncertainty intervals
frequentist: bootstrap the original model fit and save the repeated coefficient estimates (done by R rms package bootcov function); compute bootstrap confidence intervals for REVs; see Section 5.2 and example below
Bayesian: compute REV for each posterior draw of \(\beta\) and from the repeated REVs compute highest posterior density uncertainty intervals
See Luchman (2014) for a mice review of log-likelihood-based relative importance measures.
Example
Ordinal regression model (proportional odds model for continuous Y)
Allow for quadratic effects for two predictors which are allowed to interact
Three other predictors that are irrelevant are modeled linearly
Bootstrap
Through uncertainty intervals for REVs show that when the sample size is inadequate for fitting an accurate reproducible model it’s also inadequate for measuring variable importance
Plots ratios of partial \(\chi^2\) to total model LR \(\chi^2\). ascending is really a misnomer.
Figure 5.1: Relative LR \(\chi^2\) explained. Interaction effects are added to main effects.
Code
g <-bootcov(f, B=300)
Did not converge in 15 iterations
Code
plot(rexVar(g, data=d))
Figure 5.2: Relative explained variation due to each predictor. Interaction effects are added to main effects. Intervals are 0.95 bootstrap percentile confidence intervals.
5.2 The Bootstrap
If know population model, use simulation or analytic derivations to study behavior of statistical estimator
Suppose \(Y\) has a cumulative dist. fctn. \(F(y) = \Pr\{Y \leq y\}\)
We have sample of size \(n\) from \(F(y)\), \(Y_{1}, Y_{2}, \ldots, Y_{n}\)
Steps:
Repeatedly simulate sample of size \(n\) from \(F\)
Compute statistic of interest
Study behavior over \(B\) repetitions
Example: 1000 samples, 1000 sample medians, compute their sample variance
Figure 5.3: Empirical and population cumulative distribution function
\(F_{n}\) corresponds to density function placing probability \(\frac{1}{n}\) at each observed data point (\(\frac{k}{n}\) if point duplicated \(k\) times)
Pretend that \(F \equiv F_{n}\)
Sampling from \(F_{n} \equiv\) sampling with replacement from observed data \(Y_{1},\ldots,Y_{n}\)
Large \(n\)\(\rightarrow\) selects \(1-e^{-1} \approx 0.632\) of original data points in each bootstrap sample at least once
Some observations not selected, others selected more than once
Efron’s bootstrap\(\rightarrow\) general-purpose technique for estimating properties of estimators without assuming or knowing distribution of data \(F\)
Take \(B\) samples of size \(n\) with replacement, choose \(B\) so that summary measure of individual statistics \(\approx\) summary if \(B=\infty\)
Bootstrap based on distribution of observed differences between a resampled parameter estimate and the original estimate telling us about the distribution of unobservable differences between the original estimate and the unknown parameter
JKL
Example: Data \((1,5,6,7,8,9)\), obtain 0.80 confidence interval for population median, and estimate of population expected value of sample median (only to estimate the bias in the original estimate of the median).
M
Code
spar(ps=9, mfrow=c(1,2))options(digits=3)y <-c(2,5,6,7,8,9,10,11,12,13,14,19,20,21)y <-c(1,5,6,7,8,9)set.seed(17)n <-length(y)n2 <- n/2n21 <- n2+1B <-400M <-double(B)plot(0, 0, xlim=c(0,B), ylim=c(3,9),xlab="Bootstrap Samples Used",ylab="Mean and 0.1, 0.9 Quantiles", type="n")for(i in1:B) { s <-sample(1:n, n, replace=T) x <-sort(y[s]) m <- .5*(x[n2]+x[n21]) M[i] <- mpoints(i, mean(M[1:i]), pch=46)if(i>=10) { q <-quantile(M[1:i], c(.1,.9))points(i, q[1], pch=46, col='blue')points(i, q[2], pch=46, col='blue') }}table(M)
Figure 5.4: Estimating properties of sample median using the bootstrap
First 20 samples:
Bootstrap Sample
Sample Median
1 5 5 7 8 9
6.0
1 1 5 7 9 9
6.0
6 7 7 8 9 9
7.5
1 1 5 6 8 9
5.5
1 6 7 7 8 8
7.0
1 5 6 8 8 9
7.0
1 6 8 8 9 9
8.0
5 5 6 7 8 9
6.5
1 5 6 7 7 8
6.5
1 5 6 8 9 9
7.0
1 5 7 7 8 9
7.0
1 5 6 6 7 8
6.0
1 6 6 7 8 9
6.5
5 6 7 7 8 9
7.0
1 5 6 8 8 8
7.0
1 1 6 6 7 8
6.0
5 5 5 8 8 9
6.5
5 6 6 6 7 7
6.0
1 5 7 9 9 9
8.0
1 1 5 5 5 7
5.0
Histogram tells us whether we can assume normality for the bootstrap medians or need to use quantiles of medians to construct C.L.
Need high \(B\) for quantiles, low for variance (but see Booth & Sarkar (1998))
See Bradley Efron & Narasimhan (2020) for useful information about bootstrap confidence intervals and the latest R functions
N
5.3 Model Validation
5.3.1 Introduction
External validation (best: another country at another time); also validates sampling, measurements
Internal
apparent (evaluate fit on same data used to create fit)
data splitting
cross-validation
bootstrap: get overfitting-corrected accuracy index
Best way to make model fit data well is to discard much of the data
Predictions on another dataset will be inaccurate
Need unbiased assessment of predictive accuracy
OP
Working definition of external validation: Validation of a prediction tool on a sample that was not available at publication time.
Alternate: Validation of a prediction tool by an independent research team.
One suggested hierarchy of the quality of various validation methods is as follows, ordered from worst to best.
Attempting several validations (internal or external) and reporting only the one that “worked”
Reporting apparent performance on the training dataset (no validation)
Reporting predictive accuracy on an undersized independent test sample
Internal validation using data splitting where at least one of the training and test samples is not huge and the investigator is not aware of the arbitrariness of variable selection done on a single sample
Strong internal validation using 100 repeats of 10-fold cross-validation or several hundred bootstrap resamples, repeating all analysis steps involving \(Y\) afresh at each re-sample and the arbitrariness of selected “important variables” is reported (if variable selection is used)
External validation on a large test sample, done by the original research team
Re-analysis by an independent research team using strong internal validation of the original dataset
External validation using new test data, done by an independent research team
External validation using new test data generated using different instruments/technology, done by an independent research team
QR
Some points to consider:
Unless both sample sizes are huge, external validation can be low precision
External validation can be costly and slow and may result in disappointment that would have been revealed earlier with rigorous internal validation
External validation is sometimes gamed; researchers disappointed in the validation sometimes ask for a “do over”; resampling validation is harder to game as long as all analytical steps using \(Y\) are repeated each time.
Instead of external validation to determine model applicability at a different time or place, and being disappointed if the model does not work in that setting, consider building a unified model containing time and place as predictors
When the model was fully pre-specified, external validation tests the model
But when the model was fitted using machine learning, feature screening, variable selection, or model selection, the model developed using training data is usually only an example of a model, and the test sample validation could be called an example validation
When resampling is used to repeat all modeling steps for each resample, rigorous internal validation tests the process used to develop the model and happens to also provide a high-precision estimate of the likely future performance of the “final” model developed using that process, properly penalizing for model uncertainty.
Resampling also reveals the volatility of the model selection process
Collins et al. (2016) estimate that a typical sample size needed for externally validating a time-to-event model is 200 events.
5.3.2 Which Quantities Should Be Used in Validation?
OLS: \(R^2\) is one good measure for quantifying drop-off in predictive ability
Example: \(n=10, p=9\), apparent \(R^{2}=1\) but \(R^2\) will be close to zero on new subjects
Example: \(n=20, p=10\), apparent \(R^{2}=.9\), \(R^2\) on new data 0.7, \(R^{2}_{adj} = 0.79\)
Adjusted \(R^2\) solves much of the bias problem assuming \(p\) in its formula is the largest number of parameters ever examined against \(Y\)
Few other adjusted indexes exist
Also need to validate models with phantom d.f.
Cross-validation or bootstrap can provide unbiased estimate of any index; bootstrap has higher precision
Two main types of quantities to validate
Calibration or reliability: ability to make unbiased estimates of response (\(\hat{Y}\) vs. \(Y\))
Discrimination: ability to separate responses OLS: \(R^2\); \(g\)-index; binary logistic model: ROC area, equivalent to rank correlation between predicted probability of event and 0/1 event
Unbiased validation nearly always necessary, to detect overfitting
UV
5.3.3 Data-Splitting
Split data into training and test sets
Interesting to compare index of accuracy in training and test
Freeze parameters from training
Make sure you allow \(R^{2} = 1-SSE/SST\) for test sample to be \(<0\)
Don’t compute ordinary \(R^2\) on \(X\hat{\beta}\) vs. \(Y\); this allows for linear recalibration \(aX\hat{\beta} + b\) vs. \(Y\)
Test sample must be large enough to obtain very accurate assessment of accuracy
Training sample is what’s left
Example: overall sample \(n=300\), training sample \(n=200\), develop model, freeze \(\hat{\beta}\), predict on test sample (\(n=100\)), \(R^{2} = 1 -
\frac{\sum(Y_{i}-X_{i}\hat{\beta})^{2}}{\sum(Y_{i}-\bar{Y})^{2}}\).
Develop model on \(\frac{9}{10}\), including any variable selection, pre-testing, etc.
Freeze coefficients, evaluate on \(\frac{1}{10}\)
Average \(R^2\) over 10 reps
Drawbacks:
Choice of number of groups and repetitions
Doesn’t show full variability of var. selection
Does not validate full model
Lower precision than bootstrap
Need to do 50 repeats of 10-fold cross-validation to ensure adequate precision
Randomization method
Randomly permute \(Y\)
Optimism = performance of fitted model compared to what expect by chance
ZAB
5.3.5 Validation Using the Bootstrap
Estimate optimism of final whole sample fit without holding out data
From original \(X\) and \(Y\) select sample of size \(n\) with replacement
Derive model from bootstrap sample
Apply to original sample
Simple bootstrap uses average of indexes computed on original sample
Estimated optimism = difference in indexes
Repeat about \(B=100\) times, get average expected optimism
Subtract average optimism from apparent index in final model
Example: \(n=1000\), have developed a final model that is hopefully ready to publish. Call estimates from this final model \(\hat{\beta}\).
final model has apparent \(R^2\) (\(R^{2}_{app}\)) =0.4
how inflated is \(R^{2}_{app}\)?
get resamples of size 1000 with replacement from original 1000
for each resample compute \(R^{2}_{boot}\) = apparent \(R^2\) in bootstrap sample
freeze these coefficients (call them \(\hat{\beta}_{boot}\)), apply to original (whole) sample \((X_{orig}, Y_{orig})\) to get \(R^{2}_{orig} = R^{2}(X_{orig}\hat{\beta}_{boot}, Y_{orig})\)
optimism = \(R^{2}_{boot} - R^{2}_{orig}\)
average over \(B=100\) optimisms to get \(\overline{optimism}\)
Stepwise modeling often reduces optimism but this is not offset by loss of information from deleting marginal var.
G
Method
Apparent Rank Correlation of Predicted vs. Observed
Over-optimism
Bias-Corrected Correlation
Full Model
0.50
0.06
0.44
Stepwise Model
0.47
0.05
0.42
In this example, stepwise modeling lost a possible \(0.50 - 0.47 = 0.03\) predictive discrimination. The full model fit will especially be an improvement when
The stepwise selection deleted several variables which were almost significant.
These marginal variables have some real predictive value, even if it’s slight.
There is no small set of extremely dominant variables that would be easily found by stepwise selection.
H
Other issues:
See van Houwelingen & le Cessie (1990) for many interesting ideas
Faraway (1992) shows how bootstrap is used to penalize for choosing transformations for \(Y\), outlier and influence checking, variable selection, etc. simultaneously
Brownstone (1988), p. 74 feels that “theoretical statisticians have been unable to analyze the sampling properties of (usual multi-step modeling strategies) under realistic conditions” and concludes that the modeling strategy must be completely specified and then bootstrapped to get consistent estimates of variances and other sampling properties
See Blettner & Sauerbrei (1993) and Chatfield (1995) for more interesting examples of problems resulting from data-driven analyses.
I
5.3.6 Multiple Imputation and Resampling-Based Model Validation
When there are missing data, especially missing \(X\), there are simple approaches that are sometimes used for strong internal validation, including
When the fraction of incomplete observations is small, fit the model on complete cases and validate it with the bootstrap or cross-validation.
When the fraction of incomplete observations is moderately small, do single imputation and validate the model the ordinary way as in the last bullet.
For either solution, still use multiple imputation for coefficient estimation, confidence intervals, and statistical tests.
A more general approach is needed.
Spirit of multiple imputation: average the regression coefficients over \(k\) model fits for \(k\) completed datasets
For validation: do a bootstrap validation separately for each model fit on each completed dataset
Average the \(k\) estimates of model performance
For each fit use \(B\) bootstrap resamples for a total of \(kB\) iterations
May be OK to choose lower value of \(B\) than usual because of the averaging over \(k\) validations
For \(k=20\) consider using \(B=100\), with a lower \(B\) for very large dataset
Procedure is implemented using the Hmiscfit.mult.impute function (for Hmisc 5.0-1 or later) and the rmsprocessMI function (for rms 6.6-0 or later)
Example: Simulate data making two predictors missing completely at random for \(\frac{1}{5}\) of the observations
Ordinary linear model
5 noise variables added to create more overfitting
ols(formula = y ~ age + sex + x1 + x2 + x3 + x4 + x5, data = d)
Frequencies of Missing Values Due to Each Variable
y age sex x1 x2 x3 x4 x5
0 51 41 0 0 0 0 0
Model Likelihood
Ratio Test
Discrimination
Indexes
Obs 118
LR χ2 57.26
R2 0.384
σ 1.0919
d.f. 7
R2adj 0.345
d.f. 110
Pr(>χ2) 0.0000
g 0.935
Residuals
Min 1Q Median 3Q Max
-2.95513 -0.63703 0.06559 0.64790 2.55598
β
S.E.
t
Pr(>|t|)
Intercept
-2.5095
0.7361
-3.41
0.0009
age
0.0787
0.0109
7.25
<0.0001
sex=m
-0.5208
0.2062
-2.53
0.0130
x1
-0.2570
0.3390
-0.76
0.4501
x2
0.4045
0.3322
1.22
0.2260
x3
-0.1084
0.3652
-0.30
0.7671
x4
-0.4420
0.3429
-1.29
0.2001
x5
-0.8222
0.3499
-2.35
0.0206
Fraction of missing data is 0.41 so let’s run 40 imputations
Code
a <-aregImpute(~ y + age + sex + x1 + x2 + x3 + x4 + x5,data=d, n.impute=40, pr=FALSE) # 2s
For each imputation, complete the dataset using predictive mean matching
Fit a linear model on that completed dataset and run bootstrap validation of both model performance metrics and a smooth calibration curve
Just do \(B=50\) resamples per imputaton because we are doing so many imputations
Fitting linear models is so fast that higher \(B\) could have been used
Specify a function to fit.mult.impute to tell it which extra analyses are needed per completed dataset
Code
g <-function(fit) list(validate =validate(fit, B=50),calibrate =calibrate(fit, B=50) )f <-fit.mult.impute(y ~ age + sex + x1 + x2 + x3 + x4 + x5, ols, a, data=d, pr=FALSE,fun=g, fitargs=list(x=TRUE, y=TRUE)) # 3sf
Linear Regression Model
fit.mult.impute(formula = y ~ age + sex + x1 + x2 + x3 + x4 +
x5, fitter = ols, xtrans = a, data = d, fun = g, pr = FALSE,
fitargs = list(x = TRUE, y = TRUE))
Model Likelihood
Ratio Test
Discrimination
Indexes
Obs 200
LR χ2 78.74
R2 0.325
σ 1.0623
d.f. 7
R2adj 0.300
d.f. 192
Pr(>χ2) 0.0000
g 0.814
Residuals
Min 1Q Median 3Q Max
-2.89841 -0.76497 -0.05365 0.68108 2.68169
β
S.E.
t
Pr(>|t|)
Intercept
-2.4529
0.5472
-4.48
<0.0001
age
0.0710
0.0089
8.01
<0.0001
sex=m
-0.5044
0.1824
-2.77
0.0062
x1
-0.0454
0.2779
-0.16
0.8704
x2
0.4237
0.2560
1.66
0.0995
x3
-0.1978
0.2815
-0.70
0.4831
x4
-0.4577
0.2720
-1.68
0.0940
x5
-0.4373
0.2736
-1.60
0.1116
The numbers in the top box of the output are averaged over the imputations. So are the \(\beta\)s.
Average the bootstrapped performance measures and print
Code
processMI(f, 'validate')
Index
Original
Sample
Training
Sample
Test
Sample
Optimism
Corrected
Index
Successful
Resamples
R2
0.3249
0.3444
0.2961
0.0483
0.2766
2000
MSE
1.0837
1.039
1.1299
-0.0909
1.1746
2000
g
0.8137
0.8313
0.7882
0.0431
0.7706
2000
Intercept
0
0
0.0235
-0.0235
0.0235
2000
Slope
1
1
0.9506
0.0494
0.9506
2000
For the estimated calibration curves, processMI by default shows all the individual curves from all multiple imputations (set plotall=FALSE to suppress this)
Use an option to also show the first three individual validations followed by the average
Code
cal <-processMI(f, 'calibrate', nind=3)
n=200 Mean absolute error=0.091 Mean squared error=0.0139
0.9 Quantile of absolute error=0.186
n=200 Mean absolute error=0.108 Mean squared error=0.0171
0.9 Quantile of absolute error=0.181
n=200 Mean absolute error=0.095 Mean squared error=0.0123
0.9 Quantile of absolute error=0.19
n=200 Mean absolute error=0.12 Mean squared error=0.021
0.9 Quantile of absolute error=0.235
Run plot(cal) if you want to see a full size final average calibration
5.3.7 Validation of Data Reduction
A frequently asked question is whether or not unsupervised learning (data reduction) needs to be repeated inside an internal validation loop. See this for some excellent discussion. Here are are few thoughts.
To be safe, include fresh data reduction inside each loop
If you want to take a slight risk you can do the data reduction once on the whole dataset, because of the following
Principal components (especially minor components), variable and observation clustering, and redundancy analysis can be very unstable
They way in which they are unstable is uninformed by \(Y\), so this instability can work either for or against you
Typically we are allowed to use procedures that don’t always work in our favor
Contrast this with stepwise variable selection, which always attempts to work in our favor by maximizing \(R^2\)
5.3.8 Validation of Bayesian Models
Ideas about model validation can be much different in a Bayesian modeling context. If a prior distribution on a regression coefficient is somewhat flat, that implies that very large values of the parameter are likely. So overfitting cannot officially occur in that setting. Valuable discussions about Bayesian validation issues can be found here and here.
On the other hand, overfitting can damage out-of-sample predictive accuracy, and there are Bayesian Markov chain Monte Carlo shortcuts that provide leave-out-one cross-validation statistics on the gold standard log-likelihood metric. This is elegantly covered in this Richard McElreath video.
5.4 Bootstrapping Ranks of Predictors
Order of importance of predictors not pre-specified
Researcher interested in determining “winners” and “losers”
Bootstrap useful in documenting the difficulty of this task
Get confidence limits of the rank of each predictor in the scale of partial \(\chi^2\) - d.f.
Example using OLS
J
Code
# Use the plot method for anova, with pl=FALSE to suppress actual# plotting of chi-square - d.f. for each bootstrap repetition.# Rank the negative of the adjusted chi-squares so that a rank of# 1 is assigned to the highest. It is important to tell# plot.anova.rms not to sort the results, or every bootstrap# replication would have ranks of 1,2,3,... for the stats.require(rms)n <-300set.seed(1)d <-data.frame(x1=runif(n), x2=runif(n), x3=runif(n), x4=runif(n),x5=runif(n), x6=runif(n), x7=runif(n), x8=runif(n),x9=runif(n), x10=runif(n), x11=runif(n), x12=runif(n))d$y <-with(d, 1*x1 +2*x2 +3*x3 +4*x4 +5*x5 +6*x6 +7*x7 +8*x8 +9*x9 +10*x10 +11*x11 +12*x12 +9*rnorm(n))f <-ols(y ~ x1+x2+x3+x4+x5+x6+x7+x8+x9+x10+x11+x12, data=d)B <-1000ranks <-matrix(NA, nrow=B, ncol=12)rankvars <-function(fit)rank(plot(anova(fit), sort='none', pl=FALSE))Rank <-rankvars(f)for(i in1:B) { # 6s j <-sample(1:n, n, TRUE) bootfit <-update(f, data=d, subset=j) ranks[i,] <-rankvars(bootfit) }lim <-t(apply(ranks, 2, quantile, probs=c(.025,.975)))predictor <-factor(names(Rank), names(Rank))w <-data.frame(predictor, Rank, lower=lim[,1], upper=lim[,2])require(ggplot2)ggplot(w, aes(x=predictor, y=Rank)) +geom_point() +coord_flip() +scale_y_continuous(breaks=1:12) +geom_errorbar(aes(ymin=lim[,1], ymax=lim[,2]), width=0)
Figure 5.5: Bootstrap percentile 0.95 confidence limits for ranks of predictors in an OLS model. Ranking is on the basis of partial \(\chi^2\) minus d.f. Point estimates are original ranks
5.5 Simplifying the Final Model by Approximating It
5.5.1 Difficulties Using Full Models
Predictions are conditional on all variables, standard errors \(\uparrow\) when predict for a low-frequency category
Collinearity
Can average predictions over categories to marginalize, \(\downarrow\) s.e.
K
5.5.2 Approximating the Full Model
Full model is gold standard
Approximate it to any desired degree of accuracy
If approx. with a tree, best c-v tree will have 1 obs./node
Can use least squares to approx. model by predicting \(\hat{Y} =
X\hat{\beta}\)
When original model also fit using least squares, coef. of approx. model against \(\hat{Y} \equiv\) coef. of subset of variables fitted against \(Y\) (as in stepwise)
Model approximation still has some advantages
Uses unbiased estimate of \(\sigma\) from full fit
Stopping rule less arbitrary
Inheritance of shrinkage
If estimates from full model are \(\hat{\beta}\) and approx.
model is based on a subset \(T\) of predictors \(X\), coef. of approx.
model are \(W \hat{\beta}\), where \(W = (T'T)^{-1}T'X\)
Variance matrix of reduced coef.: \(W V W'\)
LM
5.6 How Do We Break Bad Habits?
Insist on validation of predictive models and discoveries
Show collaborators that split-sample validation is not appropriate unless the number of subjects is huge
Split more than once and see volatile results
Calculate a confidence interval for the predictive accuracy in the test dataset and show that it is very wide
Run simulation study with no real associations and show that associations are easy to find
Analyze the collaborator’s data after randomly permuting the \(Y\) vector and show some positive findings
Show that alternative explanations are easy to posit
Importance of a risk factor may disappear if 5 “unimportant” risk factors are added back to the model
Omitted main effects can explain apparent interactions
Uniqueness analysis: attempt to predict the predicted values from a model derived by data torture from all of the features not used in the model
N
5.7 Study Questions
Section 5.1
What is a general way to solve the problem of how to understand a complex model?
Why is it useful to estimate the effect of changing a continuous predictor on the outcome by varying the predictor from its \(25^{th}\) to its \(75^{th}\) percentile?
Section 5.2
Describe briefly in very general terms what the bootstrap involves.
The bootstrap may be used to estimate measures of many types of statistical quantities. Which type is being estimated when it comes to model validation?
Section 5.3
What are some disadvantages of external validation?
What would cause an independently validated \(R^2\) to be negative?
A model to predict the probability that a patient has a disease was developed using patients cared for at hospital A. Sort the following in order from the worst way to validate the predictive accuracy of a model to the best and most stringent way. You can just specify the letters corresponding to the following, in the right order.
compute the predictive accuracy of the model when applied to a large number of patients in another country
compute the accuracy on a large number of patients in hospital B that is in the same state as hospital A
compute the accuracy on 5 patients from hospital \(A\) that were collected after the original sample used in model fitting was completed
use the bootstrap
randomly split the data into two halves assuming that the dataset is not extremely large; develop the model on one half, freeze it, and evaluate the predictive accuracy on the other half
report how well the model predicted the disease status of the patients used to fit the model
compute the accuracy on patients in hospital \(A\) that were admitted after the last patient’s data used to fit the model were collected, assuming that both samples were large
Section 5.6
What is a way to do a misleading interaction analysis?
What would cause the most embarrassment to a researcher who analyzed a high-dimensional dataset to find “winning” associations?
Blettner, M., & Sauerbrei, W. (1993). Influence of model-building strategies on the results of a case-control study. Stat Med, 12, 1325–1338.
Booth, J. G., & Sarkar, S. (1998). Monte Carlo approximation of bootstrap variances. Am Statistician, 52, 354–357.
number of resamples required to estimate variances, quantiles; 800 resamples may be required to guarantee with 0.95 confidence that the relative error of a variance estimate is 0.1;Efron’s original suggestions for as low as 25 resamples were based on comparing stability of bootstrap estimates to sampling error, but small relative effects can significantly change P-values;number of bootstrap resamples
Breiman, L. (1992). The little bootstrap and other methods for dimensionality selection in regression: X-fixed prediction error. J Am Stat Assoc, 87, 738–754.
Brownstone, D. (1988). Regression strategies. Proceedings of the 20th Symposium on the Interface Between Computer Science and Statistics, 74–79.
Chatfield, C. (1995). Model uncertainty, data mining and statistical inference (with discussion). J Roy Stat Soc A, 158, 419–466.
bias by selecting model because it fits the data well; bias in standard errors;P. 420: ... need for a better balance in the literature and in statistical teaching between techniques and problem solving strategies. P. 421: It is “well known” to be “logically unsound and practically misleading” (Zhang, 1992) to make inferences as if a model is known to be true when it has, in fact, been selected from the same data to be used for estimation purposes. However, although statisticians may admit this privately (Breiman (1992) calls it a “quiet scandal”), they (we) continue to ignore the difficulties because it is not clear what else could or should be done. P. 421: Estimation errors for regression coefficients are usually smaller than errors from failing to take into account model specification. P. 422: Statisticians must stop pretending that model uncertainty does not exist and begin to find ways of coping with it. P. 426: It is indeed strange that we often admit model uncertainty by searching for a best model but then ignore this uncertainty by making inferences and predictions as if certain that the best fitting model is actually true. P. 427: The analyst needs to assess the model selection process and not just the best fitting model. P. 432: The use of subset selection methods is well known to introduce alarming biases. P. 433: ... the AIC can be highly biased in data-driven model selection situations. P. 434: Prediction intervals will generally be too narrow. In the discussion, Jamal R. M. Ameen states that a model should be (a) satisfactory in performance relative to the stated objective, (b) logically sound, (c) representative, (d) questionable and subject to on-line interrogation, (e) able to accommodate external or expert information and (f) able to convey information.
Collins, G. S., Ogundimu, E. O., & Altman, D. G. (2016). Sample size considerations for the external validation of a multivariable prognostic model: A resampling study. Stat Med, 35(2), 214–226. https://doi.org/10.1002/sim.6787
Copas, J. B. (1987). Cross-validation shrinkage of regression predictors. J Roy Stat Soc B, 49, 175–183.
Efron, B. (1983). Estimating the error rate of a prediction rule: Improvement on cross-validation. J Am Stat Assoc, 78, 316–331.
suggested need at least 200 models to get an average that is adequate, i.e., 20 repeats of 10-fold cv
Efron, Bradley, & Narasimhan, B. (2020). The Automatic Construction of Bootstrap Confidence Intervals. Journal of Computational and Graphical Statistics, 0(0), 1–12. https://doi.org/10.1080/10618600.2020.1714633
Efron, Bradley, & Tibshirani, R. (1993). An Introduction to the Bootstrap. Chapman and Hall.
Efron, Bradley, & Tibshirani, R. (1997). Improvements on cross-validation: The .632+ bootstrap method. J Am Stat Assoc, 92, 548–560.
Faraway, J. J. (1992). The cost of data analysis. J Comp Graph Stat, 1, 213–229.
Freedman, D., Navidi, W., & Peters, S. (1988). On the Impact of Variable Selection in Fitting Regression Equations (pp. 1–16). Springer-Verlag.
Karvanen, J., & Harrell, F. E. (2009). Visualizing covariates in proportional hazards model. Stat Med, 28, 1957–1966.
Luchman, J. N. (2014). Relative Importance Analysis With Multicategory Dependent Variables:: An Extension and Review of Best Practices. Organizational Research Methods, 17(4), 452–471. https://doi.org/10.1177/1094428114544509
Measures based on pseudo R^2 and considering all possible subsets of covariates. Good background section with review of pseudo R^2 measures.
Roecker, E. B. (1991). Prediction error and its estimation for subset-selected models. Technometrics, 33, 459–468.
Sauerbrei, W., & Schumacher, M. (1992). A bootstrap resampling procedure for model building: Application to the Cox regression model. Stat Med, 11, 2093–2109.
Shao, J. (1993). Linear model selection by cross-validation. J Am Stat Assoc, 88, 486–494.
Spiegelhalter, D. J. (1986). Probabilistic prediction in patient management and clinical trials. Stat Med, 5, 421–433. https://doi.org/10.1002/sim.4780050506
z-test for calibration inaccuracy (implemented in Stata, and R Hmisc package’s val.prob function)
Van Calster, B., Nieboer, D., Vergouwe, Y., De Cock, B., Pencina, M. J., & Steyerberg, E. W. (2016). A calibration hierarchy for risk models was defined: From utopia to empirical data. J Clin Epi, 74, 167–176. https://doi.org/10.1016/j.jclinepi.2015.12.005
van Houwelingen, J. C., & le Cessie, S. (1990). Predictive value of statistical models. Stat Med, 9, 1303–1325.
Wu, C. F. J. (1986). Jackknife, bootstrap and other resampling methods in regression analysis. Ann Stat, 14(4), 1261–1350.
Source Code
```{r include=FALSE}options(qproject='rms', prType='html')require(Hmisc)require(qreport)getRs('qbookfun.r')hookaddcap()knitr::set_alias(w = 'fig.width', h = 'fig.height', cap = 'fig.cap', scap ='fig.scap')```# Describing, Resampling, Validating, and Simplifying the Model {#sec-val}## Describing the Fitted Model {#sec-val-describe-model}`r mrg(sound("v-1"))`### Interpreting Effects* Regression coefficients if 1 d.f. per factor, no interaction `r ipacue()`* **Not** standardized regression coefficients* Many programs print meaningless estimates such as effect of increasing age$^2$ by one unit, holding age constant* Need to account for nonlinearity, interaction, and use meaningful ranges* For monotonic relationships, estimate $X\hat{\beta}$ at quartiles of continuous variables, separately for various levels of interacting factors* Subtract estimates, anti-log, e.g., to get inter-quartile-range `r ipacue()` odds or hazards ratios. Base C.L. on s.e. of difference. See @fig-coxcase-summary.* Partial effect plot: Plot effect of each predictor on $X\beta$ or some transformation. See @fig-coxcase-shapes. See also @kar09vis. Fox and Weisberg have [other excellent approaches](https://cran.r-project.org/web/packages/effects/vignettes/predictor-effects-gallery.pdf) to displaying predictor effects* Nomogram. See @fig-coxcase-nomogram* Use regression tree to approximate the full model#### Contrasts {#sec-val-contrast}Forming contrasts through differences in predicted values is the most general and easy way to estimate effects of interest in a model. This is because* One needs only to specify two or more sets of predictor settings to contrast* Predictor coding, nonlinear terms, and interactions are automatically taken into account* Unspecified predictors can default to reference values such as medians for continuous predictors and modes for categorical ones. The values used for unspecified predictors only matter if they interact with predictors that are changing in the contrast.* Approximate methods are available for getting simultaneous confidence intervals over a whole sequence of contrasts (e.g., simultaneous confidence bands for an entire range of ages) when contrasts produce more than one estimate* The approach extends to any order of differences. E.g., interaction contrasts are formed with double differences.* Contrasts allow assessment of linearity over a given predictor range in a spline fit* Contrasts allow assessment of "partial interactions", i.e., whether there is an interaction between one predictor and a subset of the levels of another predictor* The R `predict` function provides the $X$ (design matrix) that goes along with each requested prediction, and simple matrix algebra provides needed covariance matrices once differences in design matrices are formedOn the latter point, a single difference contrast such as one comparing 30 year old males with 40 year old males would generate two design matrices $X_{1}, X{2}$, and the contrast $X_{1}\beta - X_{2}\beta = (X_{1} - X_{2})\beta$.Consider a series of examples in which predicted values from a fitted model are symbolized by a function $f$ of the predictor settings, with continuous predictors $A$ and $B$ and a categorical predictor $C$ with levels $i, j, k$. Let the reference (detault) values of the predictors be $a$ (median of $A$), $b$ (median of $B$), and $c$ (modal (most frequent) category of $C$). [The term "average slope" used below will be **the** slope if the predictor has a linear effect, otherwise it is the average slope over the indicated interval. See [here](https://stats.stackexchange.com/questions/639430) for more information.]{.aside}##### Single Difference Contrasts* Average slope of $A$ over $A \in [1, 2]$: $f(2, b, c) - f(1, b, c)$* Difference between categories $j$ and $k$ in $C$: $f(a, b, j) - f(a, b, k)$* Effect of an entire range of predictor values against a single reference value: $f(1:10, b, c) - f(5, b, c)$ (a simultaneous confidence band would be useful here)##### Double Difference Contrasts* Linearity in $A$ over $A \in [1, 3]$: $f(3, b, c) - f(2, b, c) - [f(2, b, c) - f(1, b, c)] = f(3, b, c) - 2f(2, b, c) + f(1, b, c)$* Interaction between $A$ and $C$ with respect to only two levels of $C$: $f(2, b, j) - f(1, b, j) - [f(2, b, k) - f(1, b, k)]$* Interaction between $A$ and $B$ when $A$ varies from 1 to 3 and $B$ varies from 5 to 6: $f(3, 5, c) - f(1, 5, c) - [f(3, 6, c) - f(1, 6, c)]$* Effect of $C=i$ vs. $C=j$ over a regular grid of values of $B$: $f(a, 5:15, i) - f(a, 5:15, j)$ (with possible simultaneous band; this is often a better way of getting hazard ratios in a Cox proportional hazards model, as the user has control of the reference value)##### Triple Difference Contrasts* Third-order interactions* Assessment of nonlinearity in the $A\times B$ interaction (differences in double differences)The R `rms` package [`contrast` function](https://www.rdocumentation.org/packages/rms/versions/6.7-1/topics/contrast.rms) (full name `contrast.rms`) provides predicted values, differences, and double difference contrasts. The latter is specified by providing four lists of predictor settings.### Indexes of Model Performance {#sec-val-index}`r mrg(sound("v-2"))`#### Error Measures* Central tendency of prediction errors `r ipacue()` + Mean absolute prediction error: mean $|Y - \hat{Y}|$ + Mean squared prediction error - Binary $Y$: Brier score (quadratic proper scoring rule) + Logarithmic proper scoring rule (avg. log-likelihood)* Discrimination measures `r ipacue()` + Pure discrimination: rank correlation of $(\hat{Y}, Y)$ - Spearman $\rho$, Kendall $\tau$, Somers' $D_{xy}$ - $Y$ binary $\rightarrow$ $D_{xy} = 2\times (C - \frac{1}{2})$ <br> $C$ = concordance probability = area under receiver operating characteristic curve $\propto$ Wilcoxon-Mann-Whitney statistic + Mostly discrimination: $R^{2}$ - $R^{2}_{\mathrm{adj}}$---overfitting corrected if model pre-specified + Brier score can be decomposed into discrimination and calibration components + Discrimination measures based on variation in $\hat{Y}$ - regression sum of squares - $g$--index - see [here](https://fharrell.com/post/addvalue) for related information * Calibration measures `r ipacue()` + calibration--in--the--large: average $\hat{Y}$ vs. average $Y$ + high-resolution calibration curve (calibration--in--the--small). See @fig-titanic-calibrate. + calibration slope and intercept + maximum absolute calibration error + mean absolute calibration error + 0.9 quantile of calibration errorSee @cal16cal for a nice discussion ofdifferent levels of calibration stringency and their relationship tolikelihood of errors in decision making.**$g$--Index**`r mrg(sound("v-3"))`* Based on Gini's mean difference `r ipacue()` + mean over all possible $i \neq j$ of $|Z_{i} - Z_{j}|$ + interpretable, robust, highly efficient measure of variation* $g =$ Gini's mean difference of $X_{i}\hat{\beta} = \hat{Y}$* Example: $Y=$ systolic blood pressure; $g = 11$mmHg is typical difference in $\hat{Y}$* Independent of censoring etc.* For models in which anti-log of difference in $\hat{Y}$ `r ipacue()` represent meaningful ratios (odds ratios, hazard ratios, ratio of medians):<br> $g_{r} = \exp(g)$* For models in which $\hat{Y}$ can be turned into a probability estimate (e.g., logistic regression):<br> $g_{p} =$ Gini's mean difference of $\hat{P}$* These $g$--indexes represent e.g. "typical" odds ratios, "typical" risk differences* Can define partial $g$<!-- NEW 2023-09-16 -->### Relative Explained Variation {#sec-val-rev}* In OLS linear models $R^2$ is an excellent measure of explained variation in $Y$* Fraction of variance of $Y$ that is explained by $X$* SSR = $\sum_{i=1}^{n} (\hat{Y}_{i} - \bar{\hat{Y}})^2$ = regression sum of squares + proportional to variance of $\hat{Y}$* SSE = $\sum_{i=1}^{n} (Y_{i} - \hat{Y}_{i})^2$ = error sum of squares = residual sum of squares + proportional to error variance* SST = $\sum_{i=1}^{n} (Y_{i} - \bar{Y})^2$ = total sum of squares + proportion to variance of $Y$* SST = SSR + SSE* $R^{2} =$ SSR/SST = 1 - SSE/SST* For each part of the model can compute its partial SSR = loss in SSR after refitting the model upon removing the part of the model being assessed* Partial $R^{2}$ = (partial SSR)/SST* Relative partial $R^{2} = \frac{\text{partial } R^{2}}{R^{2}} =$ partial SSR divided by full model SSR* Relative partial $R^{2}$ = portion of **explainable** (by $X$) variation in $Y$ that is explained by one or more terms in the model* Relative explained variation: REV* REVs do not need to sum to 1.0 + collinearities + computing REVs for overlapping model terms* When a predictor interacts with another predictor, the predictor's REV is taken from the combined predictive information in + the predictor main effect (including any nonlinear terms) + any interaction terms involving that effect#### Extension to Other Models* For models fitted using maximum likelihood estimation, can compute the partial likelihood ratio $\chi^2$ statistic for each set and divide that by the total model LR $\chi^2$* See also @sec-mle and the example below* This is an excellent measure but requires more computation and doesn't extend to models not having a likelihood or to Bayesian modeling* Need a general purpose REV measure that works on virtually all models using all estimation techniques* Letting $\hat{Y} = X\hat{\beta}$ be the estimated linear predictor from the model* Use the same design matrix $X$ used to get $\hat{Y}$* Predict $\hat{Y}$ from $X$ using OLS* The $R^{2}$ from this model is 1.0* SSE = 0; ignore this and take variance-covariance matrix as $(X'X)^{-1}$ since SSR is used only on a relative basis* REV = $\frac{\text{partial SSR }}{\text{SSR}}$* Identical to relative $R^2$ if original model was OLS* Easily lends itself to uncertainty intervals + frequentist: bootstrap the original model fit and save the repeated coefficient estimates (done by R `rms` package `bootcov` function); compute bootstrap confidence intervals for REVs; see @sec-val-boot and example below + Bayesian: compute REV for each posterior draw of $\beta$ and from the repeated REVs compute highest posterior density uncertainty intervals* Implemented in R `rms` package [`rexVar` function](https://github.com/harrelfe/rms/blob/master/R/rexVar.r)See @luc14rel for a mice review of log-likelihood-based relative importance measures.#### Example* Ordinal regression model (proportional odds model for continuous Y)* Allow for quadratic effects for two predictors which are allowed to interact* Three other predictors that are irrelevant are modeled linearly* Bootstrap* Through uncertainty intervals for REVs show that when the sample size is inadequate for fitting an accurate reproducible model it's also inadequate for measuring variable importance```{r}require(rms)options(prType='html') #<1>n <-60set.seed(3)x1 <-rnorm(n)x2 <-rnorm(n)x3 <-rnorm(n)x4 <-rnorm(n)x5 <-rnorm(n)y <- x1 + x2 +rnorm(n)d <-data.frame(x1, x2, x3, x4, y)f <-orm(y ~pol(x1, 2) *pol(x2, 2) + x3 + x4 + x5,x=TRUE, y=TRUE) #<2>a <-anova(f, test='LR') #<3>a```1. Makes `print` methods output html for prettier printing2. `x=TRUE, y=TRUE` are needed for later bootstrapping3. Computes likelihood ratio $\chi^2$ test statistics for all partial effects in the model plus for the whole model```{r}#| label: fig-val-lr#| fig-cap: Relative LR $\chi^2$ explained. Interaction effects are added to main effects.#| fig-height: 3.5plot(a, what='proportion chisq', sort='ascending') #<1>```1. Plots ratios of partial $\chi^2$ to total model LR $\chi^2$. `ascending` is really a misnomer.```{r}#| label: fig-val-rex#| fig-cap: Relative explained variation due to each predictor. Interaction effects are added to main effects. Intervals are 0.95 bootstrap percentile confidence intervals.#| fig-height: 3.5g <-bootcov(f, B=300)plot(rexVar(g, data=d))```## The Bootstrap {#sec-val-boot}`r mrg(sound("v-4"))`* If know population model, use simulation or analytic derivations `r ipacue()` to study behavior of statistical estimator* Suppose $Y$ has a cumulative dist. fctn. $F(y) = \Pr\{Y \leq y\}$* We have sample of size $n$ from $F(y)$, <br> $Y_{1}, Y_{2}, \ldots, Y_{n}$* Steps: 1. Repeatedly simulate sample of size $n$ from $F$ 1. Compute statistic of interest 1. Study behavior over $B$ repetitions* Example: 1000 samples, 1000 sample medians, compute their sample variance `r ipacue()`* $F$ unknown $\rightarrow$ estimate by empirical dist. fctn.$$F_{n}(y) = \frac{1}{n}\sum_{i=1}^{n} [Y_{i} \leq y].$$* Example: sample of size $n=30$ from a normal distribution with mean 100 and SD 10```{r cap='Empirical and population cumulative distribution function'}#| label: fig-val-cdfspar()set.seed(6)x <- rnorm(30, 100, 20)xs <- seq(50, 150, length=150)cdf <- pnorm(xs, 100, 20)plot(xs, cdf, type='l', ylim=c(0,1), xlab=expression(x), ylab=expression(paste("Pr(", X <= x, ")")))lines(ecdf(x), cex=.5)```* $F_{n}$ corresponds to density function placing probability `r ipacue()` $\frac{1}{n}$ at each observed data point ($\frac{k}{n}$ if point duplicated $k$ times)* Pretend that $F \equiv F_{n}$* Sampling from $F_{n} \equiv$ sampling with replacement from observed data $Y_{1},\ldots,Y_{n}$* Large $n$ $\rightarrow$ selects $1-e^{-1} \approx 0.632$ of original data points in each bootstrap sample at least once* Some observations not selected, others selected more than once* Efron's _bootstrap_ $\rightarrow$ `r ipacue()` general-purpose technique for estimating properties of estimators without assuming or knowing distribution of data $F$* Take $B$ samples of size $n$ with replacement, choose $B$ `r ipacue()` so that summary measure of individual statistics $\approx$ summary if $B=\infty$* Bootstrap based on distribution of_observed_ differences between a resampled parameter estimateand the original estimate telling us about the distribution of_unobservable_ differences between the original estimate and theunknown parameterExample: Data $(1,5,6,7,8,9)$, obtain 0.80 confidence interval for `r ipacue()` population median, and estimate of population expected value of sample median (only to estimate the bias in the original estimate of the median).<!-- In 1st edition used .Random.seed <- c(1,11,4,9,61,2,29,53,34,4,16,3)--->```{r h=3, w=6, cap='Estimating properties of sample median using the bootstrap'}#| label: fig-val-bootspar(ps=9, mfrow=c(1,2))options(digits=3)y <- c(2,5,6,7,8,9,10,11,12,13,14,19,20,21)y <- c(1,5,6,7,8,9)set.seed(17)n <- length(y)n2 <- n/2n21 <- n2+1B <- 400M <- double(B)plot(0, 0, xlim=c(0,B), ylim=c(3,9), xlab="Bootstrap Samples Used", ylab="Mean and 0.1, 0.9 Quantiles", type="n")for(i in 1:B) { s <- sample(1:n, n, replace=T) x <- sort(y[s]) m <- .5*(x[n2]+x[n21]) M[i] <- m points(i, mean(M[1:i]), pch=46) if(i>=10) { q <- quantile(M[1:i], c(.1,.9)) points(i, q[1], pch=46, col='blue') points(i, q[2], pch=46, col='blue') }}table(M)hist(M, nclass=length(unique(M)), xlab="", main="")```First 20 samples:| Bootstrap Sample | | Sample Median ||-----|-----|-----|| 1 5 5 7 8 9 | | 6.0 || 1 1 5 7 9 9 | | 6.0 || 6 7 7 8 9 9 | | 7.5 || 1 1 5 6 8 9 | | 5.5 || 1 6 7 7 8 8 | | 7.0 || 1 5 6 8 8 9 | | 7.0 || 1 6 8 8 9 9 | | 8.0 || 5 5 6 7 8 9 | | 6.5 || 1 5 6 7 7 8 | | 6.5 || 1 5 6 8 9 9 | | 7.0 || 1 5 7 7 8 9 | | 7.0 || 1 5 6 6 7 8 | | 6.0 || 1 6 6 7 8 9 | | 6.5 || 5 6 7 7 8 9 | | 7.0 || 1 5 6 8 8 8 | | 7.0 || 1 1 6 6 7 8 | | 6.0 || 5 5 5 8 8 9 | | 6.5 || 5 6 6 6 7 7 | | 6.0 || 1 5 7 9 9 9 | | 8.0 || 1 1 5 5 5 7 | | 5.0 |* Histogram tells us whether we can assume normality for the bootstrap `r ipacue()` medians or need to use quantiles of medians to construct C.L.* Need high $B$ for quantiles, low for variance (but see @boo98mon)* See @efr20aut for useful information about bootstrap confidence intervals and the latest `R` functions## Model Validation {#sec-val-validation}### Introduction`r mrg(sound("v-5"))`* External validation (best: another country at another time); `r ipacue()` also validates sampling, measurements\footnote{But in many cases it is better to combine data and include country or calendar time as a predictor.}* Internal + apparent (evaluate fit on same data used to create fit) + data splitting + cross-validation + bootstrap: get overfitting-corrected accuracy index* Best way to make model fit data well is to discard much of the data `r ipacue()`* Predictions on another dataset will be inaccurate* Need unbiased assessment of predictive accuracy**Working definition of external validation**: Validation of aprediction tool on a sample that was not available at publicationtime.**Alternate**: Validation of a prediction tool by anindependent research team.One suggested hierarchy of the quality of various validation methodsis as follows, ordered from worst to best.1. Attempting several validations (internal or external) and `r ipacue()` reporting only the one that "worked"1. Reporting apparent performance on the training dataset (no validation)1. Reporting predictive accuracy on an undersized independent test sample1. Internal validation using data splitting where at least one of the training and test samples is not huge and the investigator is not aware of the arbitrariness of variable selection done on a single sample1. Strong internal validation using 100 repeats of 10-fold `r ipacue()` cross-validation or several hundred bootstrap resamples, repeating _all_ analysis steps involving $Y$ afresh at each re-sample and the arbitrariness of selected "important variables" is reported (if variable selection is used)1. External validation on a large test sample, done by the original research team1. Re-analysis by an independent research team using strong internal validation of the original dataset1. External validation using new test data, done by an independent research team1. External validation using new test data generated using different instruments/technology, done by an independent research teamSome points to consider:* Unless both sample sizes are huge, external validation can be `r ipacue()` low precision* External validation can be costly and slow and may result in disappointment that would have been revealed earlier with rigorous internal validation* External validation is sometimes _gamed_; researchers disappointed in the validation sometimes ask for a "do over"; resampling validation is harder to game as long as all analytical steps using $Y$ are repeated each time.* Instead of external validation to determine model applicability `r ipacue()` at a different time or place, and being disappointed if the model does not work in that setting, consider building a unified model containing time and place as predictors* When the model was fully pre-specified, external validation tests _the model_* But when the model was fitted using machine learning, feature screening, variable selection, or model selection, the model developed using training data is usually only an example of a model, and the test sample validation could be called an _example validation_* When resampling is used to repeat _all_ modeling steps for each resample, rigorous internal validation tests the _process_ used to develop the model and happens to also provide a high-precision estimate of the likely future performance of the "final" model developed using that process, properly penalizing for model uncertainty.* Resampling also reveals the volatility of the model selection process$\rightarrow$ See [BBR](http://hbiostat.org/bbr/reg.html#internal-vs.-external-model-validation)@col16sam estimate that a typical sample sizeneeded for externally validating a time-to-event model is 200 events.### Which Quantities Should Be Used in Validation? {#sec-val-which}`r mrg(sound("v-6"))`* OLS: $R^2$ is one good measure for quantifying drop-off in `r ipacue()` predictive ability* Example: $n=10, p=9$, apparent $R^{2}=1$ but $R^2$ will be close to zero on new subjects* Example: $n=20, p=10$, apparent $R^{2}=.9$, $R^2$ on new data 0.7, $R^{2}_{adj} = 0.79$* Adjusted $R^2$ solves much of the bias problem assuming $p$ in its formula is the largest number of parameters ever examined against $Y$* Few other adjusted indexes exist* Also need to validate models with phantom d.f.* Cross-validation or bootstrap can provide unbiased estimate of `r ipacue()` any index; bootstrap has higher precision* Two main types of quantities to validate 1. Calibration or reliability: ability to make unbiased estimates of response ($\hat{Y}$ vs. $Y$) 1. Discrimination: ability to separate responses <br> OLS: $R^2$; $g$-index; binary logistic model: ROC area, equivalent to rank correlation between predicted probability of event and 0/1 event* Unbiased validation nearly always necessary, to detect overfitting### Data-Splitting {#sec-val-split}`r mrg(sound("v-7"))`* Split data into _training_ and _test_ sets `r ipacue()`* Interesting to compare index of accuracy in training and test* Freeze parameters from training* Make sure you allow $R^{2} = 1-SSE/SST$ for test sample to be $<0$* Don't compute ordinary $R^2$ on $X\hat{\beta}$ vs. $Y$; this allows for linear recalibration $aX\hat{\beta} + b$ vs. $Y$* Test sample must be large enough to obtain very accurate assessment of accuracy `r ipacue()`* Training sample is what's left* Example: overall sample $n=300$, training sample $n=200$, develop model, freeze $\hat{\beta}$, predict on test sample ($n=100$), $R^{2} = 1 - \frac{\sum(Y_{i}-X_{i}\hat{\beta})^{2}}{\sum(Y_{i}-\bar{Y})^{2}}$.* Disadvantages of data splitting: `r ipacue()` 1. Costly in $\downarrow n$ [@roe91pre; @bre92lit] 1. Requires _decision_ to split at beginning of analysis 1. Requires larger sample held out than cross-validation 1. Results vary if split again 1. Does not validate the final model (from recombined data) 1. Not helpful in getting CL corrected for var. selection 1. Nice summary of disadvantages: @ste18val### Improvements on Data-Splitting: Resampling {#sec-val-resampling}`r mrg(sound("v-8"))`<!-- ?? TODO 3rd bullet ref pg:lrm-sw5p -->* No sacrifice in sample size `r ipacue()`* Work when modeling process automated* Bootstrap excellent for studying arbitrariness of variable selection [@sau92boo]. * Cross-validation solves many problems of data splitting [@hou90pre; @sha93lin; @wu86jac; @efr83est]* Example of $\times$-validation: `r ipacue()` 1. Split data at random into 10 tenths 1. Leave out $\frac{1}{10}$ of data at a time 1. Develop model on $\frac{9}{10}$, including any variable selection, pre-testing, etc. 1. Freeze coefficients, evaluate on $\frac{1}{10}$ 1. Average $R^2$ over 10 reps* Drawbacks: 1. Choice of number of groups and repetitions 1. Doesn't show full variability of var. selection 1. Does not validate full model 1. Lower precision than bootstrap 1. Need to do 50 repeats of 10-fold cross-validation to ensure adequate precision* Randomization method `r ipacue()` 1. Randomly permute $Y$ 1. Optimism = performance of fitted model compared to what expect by chance### Validation Using the Bootstrap {#sec-val-bootval}`r mrg(sound("v-9"))`* Estimate optimism of _final whole sample fit_ without `r ipacue()` holding out data* From original $X$ and $Y$ select sample of size $n$ with replacement* Derive model from bootstrap sample* Apply to original sample* Simple bootstrap uses average of indexes computed on original sample* Estimated optimism = difference in indexes* Repeat about $B=100$ times, get average expected optimism* Subtract average optimism from apparent index in final model* Example: $n=1000$, have developed a final model that is `r ipacue()` hopefully ready to publish. Call estimates from this final model $\hat{\beta}$. + final model has apparent $R^2$ ($R^{2}_{app}$) =0.4 + how inflated is $R^{2}_{app}$? + get resamples of size 1000 with replacement from original 1000 + for each resample compute $R^{2}_{boot}$ = apparent $R^2$ in bootstrap sample + freeze these coefficients (call them $\hat{\beta}_{boot}$), apply to original (whole) sample $(X_{orig}, Y_{orig})$ to get $R^{2}_{orig} = R^{2}(X_{orig}\hat{\beta}_{boot}, Y_{orig})$ + optimism = $R^{2}_{boot} - R^{2}_{orig}$ + average over $B=100$ optimisms to get $\overline{optimism}$ + $R^{2}_{overfitting~corrected} = R^{2}_{app} - \overline{optimism}$<!-- ?? TODO first bullet ref pg:lrm-sex-age-response-boot -->* Example: Chapter 8* Is estimating unconditional (not conditional on $X$) distribution `r ipacue()` of $R^2$, etc. [@far92cos, p. 217]* Conditional estimates would require assuming the model one is trying to validate* Efron's "$.632$" method may perform better (reduce bias further) for small $n$ [@efr83est], [@boot, p. 253], @efr97impBootstrap useful for assessing calibration in addition to discrimination:`r mrg(sound("v-10"))`* Fit $C(Y|X) = X\beta$ on bootstrap sample `r ipacue()`* Re-fit $C(Y|X) = \gamma_{0} + \gamma_{1}X\hat{\beta}$ on same data* $\hat{\gamma}_{0}=0, \hat{\gamma}_{1}=1$* Test data (original dataset): re-estimate $\gamma_{0}, \gamma_{1}$* $\hat{\gamma}_{1}<1$ if overfit, $\hat{\gamma}_{0} > 0$ to compensate* $\hat{\gamma}_{1}$ quantifies overfitting and useful for improving calibration [@spi86]* Use Efron's method to estimate optimism in $(0,1)$, estimate $(\gamma_{0}, \gamma_{1})$ by subtracting optimism from $(0,1)$* See also @cop87cro and @hou90pre, p. 1318See @fre88imp for warnings about the bootstrap, and @efr83estfor variations on the bootstrap to reduce bias.Use bootstrap to choose between full and reduced models: `r mrg(sound("v-11"))`* Bootstrap estimate of accuracy for full model `r ipacue()`* Repeat, using chosen stopping rule for each re-sample* Full fit usually outperforms reduced model[@spi86]* Stepwise modeling often reduces optimism but this is not offset by loss of information from deleting marginal var.| Method | Apparent Rank Correlation of Predicted vs. Observed | Over-optimism | Bias-Corrected Correlation ||-----|-----|-----|-----|| Full Model | 0.50 | 0.06 | 0.44 || Stepwise Model | 0.47 | 0.05 | 0.42 |In this example, stepwise modeling lost a possible $0.50 - 0.47 = 0.03$predictive discrimination.The full model fit will especially be an improvement when1. The stepwise selection deleted several variables which were `r ipacue()` almost significant.1. These marginal variables have _some_ real predictive value, even if it's slight.1. There is no small set of extremely dominant variables that would be easily found by stepwise selection.Other issues:* See @hou90pre for many interesting ideas `r ipacue()`* @far92cos shows how bootstrap is used to penalize for choosing transformations for $Y$, outlier and influence checking, variable selection, etc. simultaneously* @bro88reg, p. 74 feels that "theoretical statisticians have been unable to analyze the sampling properties of (usual multi-step modeling strategies) under realistic conditions" and concludes that the modeling strategy must be completely specified and then bootstrapped to get consistent estimates of variances and other sampling properties* See @ble93inf and @cha95mod for more interesting examples of problems resulting from data-driven analyses.### Multiple Imputation and Resampling-Based Model Validation {#sec-val-mival}When there are missing data, especially missing $X$, there are simple approaches that are sometimes used for strong internal validation, including* When the fraction of incomplete observations is small, fit the model on complete cases and validate it with the bootstrap or cross-validation.* When the fraction of incomplete observations is moderately small, do single imputation and validate the model the ordinary way as in the last bullet.* For either solution, still use multiple imputation for coefficient estimation, confidence intervals, and statistical tests.A more general approach is needed.* Spirit of multiple imputation: average the regression coefficients over $k$ model fits for $k$ completed datasets* For validation: do a bootstrap validation separately for each model fit on each completed dataset* Average the $k$ estimates of model performance* For each fit use $B$ bootstrap resamples for a total of $kB$ iterations* May be OK to choose lower value of $B$ than usual because of the averaging over $k$ validations* For $k=20$ consider using $B=100$, with a lower $B$ for very large dataset* Procedure is implemented using the `Hmisc``fit.mult.impute` function (for `Hmisc` 5.0-1 or later) and the `rms``processMI` function (for `rms` 6.6-0 or later)* Example: Simulate data making two predictors missing completely at random for $\frac{1}{5}$ of the observations* Ordinary linear model* 5 noise variables added to create more overfitting```{r}require(rms)options(prType='html')set.seed(1)n <-200d <-data.frame(age =rnorm(n, 50, 10),sex =sample(c('f', 'm'), n, TRUE),x1 =runif(n),x2 =runif(n),x3 =runif(n),x4 =runif(n),x5 =runif(n) )mis <-function(x) ifelse(runif(n) <0.2, NA, x)d <-transform(d,y =0.07* (age -50) + (sex =='f') +rnorm(n),age =mis(age),sex =mis(sex) )fracmiss <-with(d, mean(is.na(age) |is.na(sex)))# Do inefficient complete case analysisf <-ols(y ~ age + sex + x1 + x2 + x3 + x4 + x5, data=d)f ```* Fraction of missing data is `r fracmiss` so let's run 40 imputations```{r}a <-aregImpute(~ y + age + sex + x1 + x2 + x3 + x4 + x5,data=d, n.impute=40, pr=FALSE) # 2s```* For each imputation, complete the dataset using predictive mean matching* Fit a linear model on that completed dataset and run bootstrap validation of both model performance metrics and a smooth calibration curve* Just do $B=50$ resamples per imputaton because we are doing so many imputations* Fitting linear models is so fast that higher $B$ could have been used* Specify a function to `fit.mult.impute` to tell it which extra analyses are needed per completed dataset```{r}g <-function(fit) list(validate =validate(fit, B=50),calibrate =calibrate(fit, B=50) )f <-fit.mult.impute(y ~ age + sex + x1 + x2 + x3 + x4 + x5, ols, a, data=d, pr=FALSE,fun=g, fitargs=list(x=TRUE, y=TRUE)) # 3sf```* The numbers in the top box of the output are averaged over the imputations. So are the $\beta$s.* Average the bootstrapped performance measures and print```{r}processMI(f, 'validate')```* For the estimated calibration curves, `processMI` by default shows all the individual curves from all multiple imputations (set `plotall=FALSE` to suppress this)* Use an option to also show the first three individual validations followed by the average```{r}#| fig.height: 6#| fig.width: 8#| column: screen-rightcal <-processMI(f, 'calibrate', nind=3)```* Run `plot(cal)` if you want to see a full size final average calibration<!-- NEW -->### Validation of Data Reduction {#sec-val-dr}A frequently asked question is whether or not unsupervised learning (data reduction) needs to be repeated inside an internal validation loop. See [this](https://stats.stackexchange.com/questions/239898) for some excellent discussion. Here are are few thoughts.* To be safe, include fresh data reduction inside each loop* If you want to take a slight risk you can do the data reduction once on the whole dataset, because of the following* Principal components (especially minor components), variable and observation clustering, and redundancy analysis can be very unstable* They way in which they are unstable is uninformed by $Y$, so this instability can work either for or against you* Typically we are allowed to use procedures that don't always work in our favor* Contrast this with stepwise variable selection, which always attempts to work in our favor by maximizing $R^2$### Validation of Bayesian Models {#sec-val-bayes}Ideas about model validation can be much different in a Bayesian modeling context. If a prior distribution on a regression coefficient is somewhat flat, that implies that very large values of the parameter are likely. So overfitting cannot officially occur in that setting. Valuable discussions about Bayesian validation issues can be found [here](https://t.co/KfZVwGfbJb) and [here](https://t.co/BmUhOLNvEu).On the other hand, overfitting can damage out-of-sample predictive accuracy, and there are Bayesian Markov chain Monte Carlo shortcuts that provide leave-out-one cross-validation statistics on the gold standard log-likelihood metric. This is elegantly covered in [this Richard McElreath video](https://youtu.be/odGAAJDlgp8?si=F6gpXe1-h2BybBs1).## Bootstrapping Ranks of Predictors {#sec-val-bootrank}`r mrg(sound("v-12"))`* Order of importance of predictors not pre-specified `r ipacue()`* Researcher interested in determining "winners" and "losers"* Bootstrap useful in documenting the difficulty of this task* Get confidence limits of the rank of each predictor in the scale of partial $\chi^2$ - d.f.* Example using OLS```{r cap='Bootstrap percentile 0.95 confidence limits for ranks of predictors in an OLS model. Ranking is on the basis of partial $\\chi^2$ minus d.f. Point estimates are original ranks',scap='Bootstrap confidence limits for ranks of predictors'}#| label: fig-val-bootrank#| fig-height: 3# Use the plot method for anova, with pl=FALSE to suppress actual# plotting of chi-square - d.f. for each bootstrap repetition.# Rank the negative of the adjusted chi-squares so that a rank of# 1 is assigned to the highest. It is important to tell# plot.anova.rms not to sort the results, or every bootstrap# replication would have ranks of 1,2,3,... for the stats.require(rms)n <- 300set.seed(1)d <- data.frame(x1=runif(n), x2=runif(n), x3=runif(n), x4=runif(n), x5=runif(n), x6=runif(n), x7=runif(n), x8=runif(n), x9=runif(n), x10=runif(n), x11=runif(n), x12=runif(n))d$y <- with(d, 1*x1 + 2*x2 + 3*x3 + 4*x4 + 5*x5 + 6*x6 + 7*x7 + 8*x8 + 9*x9 + 10*x10 + 11*x11 + 12*x12 + 9*rnorm(n))f <- ols(y ~ x1+x2+x3+x4+x5+x6+x7+x8+x9+x10+x11+x12, data=d)B <- 1000ranks <- matrix(NA, nrow=B, ncol=12)rankvars <- function(fit) rank(plot(anova(fit), sort='none', pl=FALSE))Rank <- rankvars(f)for(i in 1:B) { # 6s j <- sample(1:n, n, TRUE) bootfit <- update(f, data=d, subset=j) ranks[i,] <- rankvars(bootfit) }lim <- t(apply(ranks, 2, quantile, probs=c(.025,.975)))predictor <- factor(names(Rank), names(Rank))w <- data.frame(predictor, Rank, lower=lim[,1], upper=lim[,2])require(ggplot2)ggplot(w, aes(x=predictor, y=Rank)) + geom_point() + coord_flip() + scale_y_continuous(breaks=1:12) + geom_errorbar(aes(ymin=lim[,1], ymax=lim[,2]), width=0)```See [this](https://fharrell.com/post/badb) and [this](https://hbiostat.org/talks/older/FHbiomarkers.pdf) for a higher-dimension example.## Simplifying the Final Model by Approximating It {#sec-val-approx}`r mrg(sound("v-13"))`### Difficulties Using Full Models* Predictions are conditional on all variables, standard errors `r ipacue()` $\uparrow$ when predict for a low-frequency category* Collinearity* Can average predictions over categories to marginalize, $\downarrow$ s.e.### Approximating the Full Model* Full model is gold standard `r ipacue()`* Approximate it to any desired degree of accuracy* If approx. with a tree, best c-v tree will have 1 obs./node* Can use least squares to approx. model by predicting $\hat{Y} = X\hat{\beta}$* When original model also fit using least squares, coef. of approx. model against $\hat{Y} \equiv$ coef. of subset of variables fitted against $Y$ (as in stepwise)* Model approximation still has some advantages `r ipacue()` 1. Uses unbiased estimate of $\sigma$ from full fit 1. Stopping rule less arbitrary 1. Inheritance of shrinkage* If estimates from full model are $\hat{\beta}$ and approx.\ model is based on a subset $T$ of predictors $X$, coef. of approx.\ model are $W \hat{\beta}$, where <br> $W = (T'T)^{-1}T'X$* Variance matrix of reduced coef.: $W V W'$## How Do We Break Bad Habits? {#sec-val-habits}`r mrg(sound("v-14"))`* Insist on validation of predictive models and discoveries `r ipacue()`* Show collaborators that split-sample validation is not appropriate unless the number of subjects is huge + Split more than once and see volatile results + Calculate a confidence interval for the predictive accuracy in the test dataset and show that it is very wide* Run simulation study with no real associations and show that associations are easy to find* Analyze the collaborator's data after randomly permuting the $Y$ vector and show some positive findings* Show that alternative explanations are easy to posit + Importance of a risk factor may disappear if 5 "unimportant" risk factors are added back to the model + Omitted main effects can explain apparent interactions + _Uniqueness analysis_: attempt to predict the predicted values from a model derived by data torture from all of the features not used in the model## Study Questions**Section 5.1**1. What is a general way to solve the problem of how to understand a complex model?1. Why is it useful to estimate the effect of changing a continuous predictor on the outcome by varying the predictor from its $25^{th}$ to its $75^{th}$ percentile?**Section 5.2**1. Describe briefly in very general terms what the bootstrap involves.1. The bootstrap may be used to estimate measures of many types of statistical quantities. Which type is being estimated when it comes to model validation?**Section 5.3**1. What are some disadvantages of external validation?1. What would cause an independently validated $R^2$ to be negative?1. A model to predict the probability that a patient has a disease was developed using patients cared for at hospital A. Sort the following in order from the worst way to validate the predictive accuracy of a model to the best and most stringent way. You can just specify the letters corresponding to the following, in the right order. a. compute the predictive accuracy of the model when applied to a large number of patients in another country a. compute the accuracy on a large number of patients in hospital B that is in the same state as hospital A a. compute the accuracy on 5 patients from hospital $A$ that were collected after the original sample used in model fitting was completed a. use the bootstrap a. randomly split the data into two halves assuming that the dataset is not extremely large; develop the model on one half, freeze it, and evaluate the predictive accuracy on the other half a. report how well the model predicted the disease status of the patients used to fit the model a. compute the accuracy on patients in hospital $A$ that were admitted after the last patient's data used to fit the model were collected, assuming that both samples were large**Section 5.6**1. What is a way to do a misleading interaction analysis?1. What would cause the most embarrassment to a researcher who analyzed a high-dimensional dataset to find "winning" associations?```{r echo=FALSE}saveCap('05')```
