Implementation of the PATH Statement

The recent PATH (Predictive Approaches to Treatment effect Heterogeneity) Statement outlines principles, criteria, and key considerations for applying predictive approaches to clinical trials to provide patient-centered evidence in support of decision making. Here challenges in implementing the PATH Statement are addressed with the GUSTO-I trial as a case study.

Author

Ewout Steyerberg

Department of Biomedical Data Sciences
Leiden University Medical Center NL
@ESteyerberg

Published

November 24, 2020

Evidence is derived from groups while most medical decisions are made for individual patients. — Kent et al, PATH statement

Heterogeneity of treatment effect (HTE) refers to the nonrandom variation in the magnitude of the absolute treatment effect (treatment benefit) across individual patients. The recent PATH (Predictive Approaches to Treatment effect Heterogeneity) Statement outlines principles, criteria, and key considerations for applying predictive approaches to clinical trials to provide patient-centered evidence in support of decision making. The focus of PATH is on modeling of HTE across individual patients.

Comments

The PATH statement lists a number of principles and guidelines. A first principle is to establish overall treatment effect. In another blog, I summarized the arguments in favor of covariate adjustment as the primary analysis of a RCT. Illustration was in the GUSTO-I trial. Here I continue that illustration, also following the blog by Frank Harrell on examining HTE.

Illustration in the GUSTO-I trial

Let’s analyze the data from 30,510 patients with an acute myocardial infarction as included in the GUSTO-I trial. In GUSTO-I, 10,348 patients were randomized to receive tissue plasminogen activator (tPA), while 20,162 were randomzied to Streptokinase (SK) and had 30-day mortality status known.

Code

load(url('http://hbiostat.org/data/gusto.rda'))
# keep only SK and tPA arms; and selected set of covariates
gusto <- upData(gusto[gusto$tx=="SK" | gusto$tx=="tPA",], 
                keep=Cs(day30, tx, age, Killip, sysbp, pulse, pmi, miloc, sex),
                                print=FALSE)
html(describe(gusto), scroll=TRUE)

gusto Descriptives

gusto

9 Variables 30510 Observations

day30

n	missing	distinct	Info	Sum	Mean	Gmd
30510	0	2	0.195	2128	0.06975	0.1298

sex: Sex

n	missing	distinct
30510	0	2

 Value        male female
 Frequency   22795   7715
 Proportion  0.747  0.253

Killip: Killip Class

n	missing	distinct
30510	0	4

 Value          I    II   III    IV
 Frequency  26007  3857   417   229
 Proportion 0.852 0.126 0.014 0.008

age

n	missing	distinct	Info	Mean	Gmd	.05	.10	.25	.50	.75	.90	.95
30510	0	5342	1	60.91	13.58	40.92	44.73	52.11	61.58	69.84	76.19	79.42

lowest : 19.0 20.8 21.0 21.0 21.4 , highest: 91.9 92.3 96.5 108.0 110.0

pulse: Heart Rate beats/min

n	missing	distinct	Info	Mean	Gmd	.05	.10	.25	.50	.75	.90	.95
30510	0	157	0.999	75.38	19.5	50	55	62	73	86	98	107

lowest : 0 1 6 9 20 , highest: 191 200 205 210 220

sysbp: Systolic Blood Pressure mmHg

n	missing	distinct	Info	Mean	Gmd	.05	.10	.25	.50	.75	.90	.95
30510	0	196	0.999	129	26.58	92.0	100.0	112.0	129.5	144.0	160.0	170.0

lowest : 0 36 40 43 46 , highest: 266 274 275 276 280

miloc: MI Location

n	missing	distinct
30510	0	3

 Value      Inferior    Other Anterior
 Frequency     17582     1062    11866
 Proportion    0.576    0.035    0.389

pmi: Previous MI

n	missing	distinct
30510	0	2

 Value         no   yes
 Frequency  25452  5058
 Proportion 0.834 0.166

tx: Tx in 3 groups

n	missing	distinct
30510	0	2

 Value         SK   tPA
 Frequency  20162 10348
 Proportion 0.661 0.339

Overall treatment effect

The primary outcome was 30-day mortality. Among the tPA group, the 30-day mortality was 653/10,348 = 6.3% vs 1475/20,162 = 7.3% in the SK group. This an absolute difference of 1.0%, or an odds ratio of 0.85 [0.78-0.94].

Code

# simple cross-table
table1(~ as.factor(day30) | tx, data=gusto, digits=2)

	SK (N=20162)	tPA (N=10348)	Overall (N=30510)
as.factor(day30)
0	18687 (92.7%)	9695 (93.7%)	28382 (93.0%)
1	1475 (7.3%)	653 (6.3%)	2128 (7.0%)

Code

# 'tpa' as 0/1 variable for tPA vs SK treatment
gusto$tx <- droplevels(gusto$tx) # Drop the SK + tPA category which has 0 obs
gusto$tpa <- as.numeric(gusto$tx) - 1 # 0/1 coding for tPA treatment
label(gusto$tpa) <- "tPA"
levels(gusto$tpa) <- c("SK", "tPA")
tab2 <- table(gusto$day30, gusto$tpa)
result <- OddsRatio(tab2, conf.level = 0.95)
names(result) <- c("Odds Ratio", "Lower CI", "Upper CI")

kable(as.data.frame(t(result))) %>% kable_styling(full_width=F, position = "left")

Odds Ratio	Lower CI	Upper CI
0.853	0.776	0.939

Code

# BinomDiffCI(x1 = events1, n1 = n1, x2 = events2, n2 = n2, ...)
CI      <- BinomDiffCI(x1 = tab2[2,1], n1 = sum(tab2[,1]), x2 = tab2[2,2], n2 = sum(tab2[,2]),
                       method = "scorecc")
colnames(CI) <- c("Absolute difference", "Lower CI", "Upper CI")

result <- round(CI, 3) # absolute difference with confidence interval
kable(as.data.frame(result)) %>% kable_styling(full_width=F, position = "left")

Absolute difference	Lower CI	Upper CI
0.01	0.004	0.016

Adjustment for baseline covariates

The unadjusted odds ratio of 0.853 is a marginal estimate. As explained in the other blog, a lot can be said in favor of conditional estimates, where we adjust for prognostically important baseline characteristics. In line with Califf 1997 and Steyerberg 2000, we consider a prediction model with 6 baseline covariates, including age and Killip class (a measure for ventricular function). Pulse rate is modeled using a linear spline with a knot at 50 beats/minute.

Code

options(prType='html')
f <- lrm(day30 ~ tpa + age + Killip + pmin(sysbp, 120) + lsp(pulse, 50) + pmi + miloc, data=gusto, x=T, maxit=99)
print(f) # coef tPA: -0.2080

Logistic Regression Model

 lrm(formula = day30 ~ tpa + age + Killip + pmin(sysbp, 120) + 
     lsp(pulse, 50) + pmi + miloc, data = gusto, x = T, maxit = 99)

	Model Likelihood Ratio Test	Discrimination Indexes	Rank Discrim. Indexes
Obs 30510	LR χ² 2991.95	R² 0.235	C 0.815
0 28382	d.f. 11	R²_11,30510 0.093	D_xy 0.631
1 2128	Pr(>χ²) <0.0001	R²_11,5938.7 0.395	γ 0.631
max \|∂log L/∂β\| 6×10^-10		Brier 0.056	τ_a 0.082

	β	S.E.	Wald Z	Pr(>\|Z\|)
Intercept	-3.0203	0.7973	-3.79	0.0002
tpa	-0.2080	0.0529	-3.93	<0.0001
age	0.0769	0.0025	31.28	<0.0001
Killip=II	0.6137	0.0589	10.42	<0.0001
Killip=III	1.1610	0.1214	9.57	<0.0001
Killip=IV	1.9213	0.1618	11.87	<0.0001
sysbp	-0.0392	0.0019	-20.33	<0.0001
pulse	-0.0242	0.0159	-1.52	0.1282
pulse'	0.0433	0.0162	2.67	0.0075
pmi=yes	0.4472	0.0562	7.96	<0.0001
miloc=Other	0.2863	0.1347	2.13	0.0335
miloc=Anterior	0.5432	0.0511	10.62	<0.0001

So, we note that the adjusted regression coefficient for tPA was -0.208. The adjusted OR = 0.812.
Let’s check for statistical interaction with

Individual covariates
The linear predictor

Code

## check for interaction
# 1. traditional approach
g <- lrm(day30 ~  tx * (age + Killip + pmin(sysbp, 120) + lsp(pulse, 50) + pmi + miloc), 
          data=gusto, maxit=100)
print(anova(g)) # tx interactions: 10 df, p=0.5720; based on LR test

	χ²	d.f.	P
Wald Statistics for `day30`
tx (Factor+Higher Order Factors)	23.48	11	0.0151
All Interactions	8.53	10	0.5772
age (Factor+Higher Order Factors)	981.65	2	<0.0001
All Interactions	1.60	1	0.2065
Killip (Factor+Higher Order Factors)	281.44	6	<0.0001
All Interactions	1.37	3	0.7131
sysbp (Factor+Higher Order Factors)	412.30	2	<0.0001
All Interactions	0.11	1	0.7369
pulse (Factor+Higher Order Factors)	235.92	4	<0.0001
All Interactions	4.22	2	0.1214
Nonlinear (Factor+Higher Order Factors)	7.37	2	0.0251
pmi (Factor+Higher Order Factors)	64.23	2	<0.0001
All Interactions	0.05	1	0.8233
miloc (Factor+Higher Order Factors)	115.23	4	<0.0001
All Interactions	2.99	2	0.2239
tx × age (Factor+Higher Order Factors)	1.60	1	0.2065
tx × Killip (Factor+Higher Order Factors)	1.37	3	0.7131
tx × sysbp (Factor+Higher Order Factors)	0.11	1	0.7369
tx × pulse (Factor+Higher Order Factors)	4.22	2	0.1214
Nonlinear	0.08	1	0.7784
Nonlinear Interaction : f(A,B) vs. AB	0.08	1	0.7784
tx × pmi (Factor+Higher Order Factors)	0.05	1	0.8233
tx × miloc (Factor+Higher Order Factors)	2.99	2	0.2239
TOTAL NONLINEAR	7.37	2	0.0251
TOTAL INTERACTION	8.53	10	0.5772
TOTAL NONLINEAR + INTERACTION	15.57	11	0.1579
TOTAL	2343.20	21	<0.0001

Code

# 2. PATH statement: linear interaction with linear predictor; baseline risk, so tx=ref, SK
lp.no.tx <- f$coefficients[1] + 0 * f$coefficients[2] + f$x[,-1] %*% f$coefficients[-(1:2)]
gusto$lp <- as.vector(lp.no.tx) # add lp to data frame
h <- lrm(day30 ~ tx * lp, data=gusto)
print(anova(h)) # tx interaction: 1 df, p=0.35; based on Wald statistics

	χ²	d.f.	P
Wald Statistics for `day30`
tx (Factor+Higher Order Factors)	16.24	2	0.0003
All Interactions	0.86	1	0.3526
lp (Factor+Higher Order Factors)	2338.37	2	<0.0001
All Interactions	0.86	1	0.3526
tx × lp (Factor+Higher Order Factors)	0.86	1	0.3526
TOTAL	2342.20	3	<0.0001

So:

The overall test for interaction with the individual covariates is far from statistically significant (p>0.5).
Similarly, the test for interaction with the linear predictor is far from statistically significant (p>0.3).

Conclusion: no interaction needed

We conclude that we may proceed by ignoring any interactions. We have no evidence against the assumption that the overall effect of treatment is applicable to all patients.
The patients vary widely in risk, as can easily be seen in the histogram below.

Code

hist(plogis(lp.no.tx), xlim=c(0,.4), main="")

We note that many patients have baseline risks (tPA==0) below 5%. Obviously their maximum benefit is bounded by this risk estimate, even if tPA, hypothetically, would reduce the risk to Null.

PATH principle: perform risk-based subgrouping

Fig 3 of the PATH Statement starts with:

Reporting RCT results stratified by a risk model is encouraged when overall trial results are positive to better understand the distribution of effects across the trial population.

How do we provide such reporting of RCT results?

We can provide estimates of relative effects and absolute benefit

By group (e.g. quarters defined by quartiles)
By baseline risk (continuous, as in the histogram above)

1a. Relative effects of treatment by risk-group

The checks for interaction were far from statistically significant in GUSTO-I. We can further illustrate the relative effects in a forest plot. Many reports from RCTs include forest plots that show relative effects by single variables, such as men vs women; young vs old age; disease subtype; etc.
The PATH statement encourages reporting by risk-based subgroup. How can such reporting be done?

Let’s do some data processing to make a better forest plot.

A PATH compatible forest plot

Let’s expand the standard forest plot for subgroup effects with risk-based subgroups.
Below subgroup effects are defined for 4 risk-based groups using cut2(lp.no.tx, g=4),
and for 3 classical subgroups (by sex, age, type of infarction).

Code

## quantiles, suggest to use quarters
groups <- cut2(lp.no.tx, g=4)
group0 <- groups[gusto$tpa==0]  # SK gropup
group1 <- groups[gusto$tpa==1]  # tPA group

rate0 <- prop.table(table(group0, gusto$day30[gusto$tpa==0]),1 )[,2]
rate1 <- prop.table(table(group1, gusto$day30[gusto$tpa==1]),1 )[,2]
ratediff <- rate0-rate1 # benefit of tPA by group

# Make a data frame for the results
data.subgroups <- as.data.frame(matrix(nrow=(4+6+1), ncol=10))
colnames(data.subgroups) <- c("tevent", "tnoevent", "cevent", "cnoevent", 
                              "name", "type", "tn", "pt", "cn", "pc")

data.subgroups[11,1:4] <- table(gusto$tpa,gusto$day30)[4:1] # overall results
# define event and non-event numbers
events1   <- table(group0, gusto$day30[gusto$tpa==0])[,2]
nevents1  <- table(group0, gusto$day30[gusto$tpa==0])[,1]
events2   <- table(group1, gusto$day30[gusto$tpa==1])[,2]
nevents2  <- table(group1, gusto$day30[gusto$tpa==1])[,1]
n1      <- events1 + nevents1
n2      <- events2 + nevents2

data.subgroups[10:7,1:4] <- cbind(events2,nevents2,events1,nevents1)

Data for classic subgroups are managed below:

Code

# Use `table`  to get the summary of cell numbers, by subgroup
# SEX
data.subgroups[5,1:4] <- table(1-gusto$day30,1-gusto$tpa, gusto$sex)[1:4]
data.subgroups[6,1:4] <- table(1-gusto$day30,1-gusto$tpa, gusto$sex)[5:8]
# AGE
data.subgroups[3,1:4] <- table(1-gusto$day30,1-gusto$tpa, gusto$age>=75)[1:4]
data.subgroups[4,1:4] <- table(1-gusto$day30,1-gusto$tpa, gusto$age>=75)[5:8]
# ANT
data.subgroups[1,1:4] <- table(1-gusto$day30,1-gusto$tpa, gusto$miloc=="Anterior")[1:4]
data.subgroups[2,1:4] <- table(1-gusto$day30,1-gusto$tpa, gusto$miloc=="Anterior")[5:8]

# Names
data.subgroups[11,5]   <- "Overall"
data.subgroups[10:7,5] <- paste("Quarter",1:4, sep=" ")
data.subgroups[5:6,5]  <- c("Male sex","Female sex")
data.subgroups[3:4,5]  <- c("Age <75","Age>=75")
data.subgroups[1:2,5]  <- c("Other MI","Anterior")

# Type of subgroup
data.subgroups[11,6]   <- ""
data.subgroups[10:7,6] <- c(rep("Risk-based subgroups", length(ratediff)))
data.subgroups[1:6,6] <- c(rep("Location",2), rep("Age",2), rep("Sex",2))

data.subgroups[,7] <- data.subgroups[,1] + data.subgroups[,2]
data.subgroups[,8] <- paste(round(100*data.subgroups[,1] / data.subgroups[,7] , 1),"%", sep="")
data.subgroups[,9] <- data.subgroups[,3] + data.subgroups[,4]
data.subgroups[,10] <- paste(round(100*data.subgroups[,3] / data.subgroups[,9] , 1),"%", sep="")

# Show the data
kable(as.data.frame((data.subgroups))) %>% kable_styling(full_width=F, position = "left")

tevent	tnoevent	cevent	cnoevent	name	type	tn	pt	cn	pc
301	6025	649	11669	Other MI	Location	6326	4.8%	12318	5.3%
352	3670	826	7018	Anterior	Location	4022	8.8%	7844	10.5%
397	8622	981	16781	Age <75	Age	9019	4.4%	17762	5.5%
256	1073	494	1906	Age>=75	Age	1329	19.3%	2400	20.6%
383	7342	888	14182	Male sex	Sex	7725	5%	15070	5.9%
270	2353	587	4505	Female sex	Sex	2623	10.3%	5092	11.5%
455	2163	1000	4009	Quarter 4	Risk-based subgroups	2618	17.4%	5009	20%
130	2479	300	4719	Quarter 3	Risk-based subgroups	2609	5%	5019	6%
49	2481	130	4967	Quarter 2	Risk-based subgroups	2530	1.9%	5097	2.6%
19	2572	45	4992	Quarter 1	Risk-based subgroups	2591	0.7%	5037	0.9%
653	1475	9695	18687	Overall		2128	30.7%	28382	34.2%

In this table tevent means #events among treated; cevent means #events among non-treated; etc

Results can be plotted with metafor functions:

Code

par(mar=c(4,4,1,2))
### fit random-effects model (use slab argument to define "study" labels)
res <- rma(ai=tevent, bi=tnoevent, ci=cevent, di=cnoevent, data=data.subgroups, measure="OR",
           slab=name, method="ML")

### set up forest plot (with 2x2 table counts added); rows argument is used
### to specify exactly in which rows the outcomes will be plotted)
forest(res, xlim=c(-8, 2.5), at=log(c(0.5, 1)), alim=c(log(0.2), log(2)), atransf=exp,
       ilab=cbind(data.subgroups$tn, data.subgroups$pt, data.subgroups$cn, data.subgroups$pc),
       ilab.xpos=c(-5,-4,-3,-2), adj=1,
       cex=.9, ylim=c(0, 19),
       rows=c(1:2, (4:5)-.5, 6:7, 10:13, 15),
       xlab="", mlab="", psize=.8, addfit=F)
# lines(x=c(-.15, -.15), y=c(0, 17)) ## could add a reference line of the overall treatment effect

text(c(-5,-4,-3,-2, 2.2), 18, c("n", "%mort", "n", "%mort", "OR    [95% CI]"), 
     font=2, adj=1, cex=.9)
text(-8, 18, c("GUSTO-I trial"), font=2, adj=0, cex=.9)
text(c(-4.5,-2.5),  19, c("tPA", "SK"), font=2, adj=1)

Code

# This can be improved

This forest plot shows the unadjusted overall effect of tPA vs SK treatment; risk-based subgroup effects; and traditional one at a time subgroup effects. The latter are to be interpreted with much caution; many false-positive findings may arise.

Q: What R function can assist trialists in their reporting of risk-based subgroups together with classic subgroups?

We can also estimate the same subgroup effects, adjusted for baseline risk.

Code

# function for adjustment
subgroup.adj <- function(data=gusto, subgroup=gusto$sex) {
  coef.unadj <- by(data, subgroup, function(x)lrm.fit(y=x$day30, x=x$tpa)$coef[2])
  var.unadj  <- by(data, subgroup, function(x)lrm.fit(y=x$day30, x=x$tpa)$var[2,2])
  coef.adj   <- by(data, subgroup, function(x)lrm.fit(y=x$day30, x=x$tpa, offset=x$lp)$coef[2])
  var.adj    <- by(data, subgroup, function(x)lrm.fit(y=x$day30, x=x$tpa, offset=x$lp)$var[2,2])
  result <- cbind(coef.unadj, coef.adj, coef.ratio=coef.adj/coef.unadj, 
                  SEunadj=sqrt(var.unadj), SEadj=sqrt(var.adj), 
                  SEratio=sqrt(var.adj)/ sqrt(var.unadj)) 
  result} # end function

options(digits=3)

Overall trial result (un)adjusted

Code

kable(as.data.frame(subgroup.adj(gusto, gusto$tpa>-1))) %>% 
  kable_styling(full_width=F, position = "left")

	coef.unadj	coef.adj	coef.ratio	SEunadj	SEadj	SEratio
TRUE	-0.159	-0.208	1.31	0.049	0.053	1.09

Risk-based subgroups

Code

kable(as.data.frame(subgroup.adj(gusto, groups))) %>% 
  kable_styling(full_width=F, position = "left")

	coef.unadj	coef.adj	coef.ratio	SEunadj	SEadj	SEratio
[-7.18,-4.01)	-0.199	-0.198	0.993	0.275	0.275	1.00
[-4.01,-3.20)	-0.282	-0.281	0.998	0.169	0.170	1.00
[-3.20,-2.36)	-0.193	-0.187	0.971	0.108	0.108	1.00
[-2.36, 4.70]	-0.170	-0.208	1.219	0.063	0.067	1.07

Classic subgroup: men vs women

Code

kable(as.data.frame(subgroup.adj(gusto, gusto$sex))) %>% 
  kable_styling(full_width=F, position = "left")

	coef.unadj	coef.adj	coef.ratio	SEunadj	SEadj	SEratio
male	-0.183	-0.237	1.29	0.063	0.068	1.08
female	-0.127	-0.164	1.29	0.078	0.085	1.10

Classic subgroup: old (>=75) vs younger age

Code

kable(as.data.frame(subgroup.adj(gusto, gusto$age>=75))) %>% 
  kable_styling(full_width=F, position = "left")

	coef.unadj	coef.adj	coef.ratio	SEunadj	SEadj	SEratio
FALSE	-0.239	-0.262	1.10	0.061	0.065	1.06
TRUE	-0.083	-0.100	1.21	0.086	0.093	1.08

The unadjusted and adjusted results are usually quite in line; only subtle differences is the estimates of the relative effects are noted. We might hypothesize that the unadjusted effect for women is confounded by the higher age of women (where higher age was associated with a somewhat weaker treatment effect); this was not confirmed.

Q: What R function can be developed that extends the unadjusted forest plots to provide subgroup effects, adjusted for baseline characteristics?

1b. Absolute benefit of treatment by risk-group

Code

# 95% CI 
CI      <- BinomDiffCI(x1 = events1, n1 = n1, x2 = events2, n2 = n2, method = "scorecc")

colnames(CI) <- c("Absolute difference", "Lower CI", "Upper CI")
rownames(CI) <- names(events1)

result <- round(CI, 3) # absolute difference with confidence interval
kable(as.data.frame(result)) %>% kable_styling(full_width=F, position = "left")

	Absolute difference	Lower CI	Upper CI
[-7.18,-4.01)	0.002	-0.003	0.006
[-4.01,-3.20)	0.006	-0.001	0.013
[-3.20,-2.36)	0.010	-0.001	0.020
[-2.36, 4.70]	0.026	0.007	0.044

So, we see a substantial difference in absolute benefit. Low risk according to the linear predictor (lp<-2.36) implies low benefit (<1%); higher risk, higher benefit (>2%). As Frank Harrell would also emphasize, the grouping in quarters might primarily be considered for illustration. A better estimation of benefit avoids grouping, and conditions on the baseline risk.

2a. Relative effects of treatment by baseline risk

The checks for interaction with the linear predictor were far from statistically significant in GUSTO-I, as shown above; supporting the assumption that 1 single adjusted, relative effect applies across baseline risk.

2b. Absolute benefit of treatment by baseline risk

Estimation of absolute benefit can follow a parametric approach, i.e. following the no interaction, main effect model that includes baseline characteristics and a treatment effect; the model f considered above for the primary analysis of the trial.
Further down we will consider relaxations of the proportionality of effect that is assumed in this model.

Code

# create baseline predictions: X-axis
xp <- seq(0.002,.5,by=0.001)
logxp0 <- log(xp/(1-xp))

# expected difference, if covariate adjusted model holds
p1exp <- plogis(logxp0) - plogis(logxp0+coef(f)[2]) # proportional effect assumed

plot(x=xp, y=p1exp, type='l', lty=2, lwd=3, xlim=c(0,.35), ylim=c(-0.007,.05), col="red",
     xlab="Baseline risk", ylab="Benefit by tPA", cex.lab=1.2, las=1, bty='l' )
# add horizontal line
lines(x=c(0,.5), y=c(0,0)) 
# distribution of predicted 
histSpike(plogis(lp.no.tx), add=T, side=1, nint=300, frac=.15) 

points(x=rate0, y=ratediff, pch=1, cex=2, lwd=2, col="blue")
arrows(x0=rate0, x1=rate0, y0=CI[,2], y1=CI[,3], angle=90, code=3,len=.1, col="blue")

legend("topleft", lty=c(2,NA), pch=c(NA,1), lwd=c(3,2), bty='n',col=c("red", "blue"), cex=1.2,
       legend=c("Expected with proportional effect", 
                "Grouped patients"))

This plot shows the benefit by tPA treatment over SK. The red line assumes a proportional effect of treatment, which may be quite reasonable here and in many other diseases. The quarters provide for a non-parametric confirmation of the benefit across baseline risk.

Relaxation of the proportional effect assumption

If we want to relax the proportional effect assumption, the blog by Frank Harrell on examining HTE provides an illustration with penalized logistic regression.

Another possible relaxation is by including interaction with the linear predictor. We consider linear interaction and a non-linear interaction (rcs, 2 df for non-linearity).
And we could try a more non-paramteric approach as in Califf 1997. There, loess smoothers were used for risk estimation the tPA (day30~lp, subset=tpa==1) and SK groups (day30~lp, subset=tpa==0).
Benefit was the differences between these 2 risk groups conditional on baseline risk.

Code

h  <- lrm(day30 ~ tpa + tpa * lp, data=gusto, eps=0.005, maxit=30)
h2 <- lrm(day30 ~ tpa + rcs(lp,3)*tpa, data=gusto, eps=0.005, maxit=99)
# loess smoothing
l0 <- loess(day30 ~ lp, data=gusto, subset=tpa==0)
l1 <- loess(day30 ~ lp, data=gusto, subset=tpa==1)

# subtract predicted risks with from without tx
p1 <- plogis(Predict(h,  tpa=0, lp = logxp0)[,3]) - 
      plogis(Predict(h,  tpa=1, lp = logxp0)[,3])
p2 <- plogis(Predict(h2, tpa=0, lp = logxp0)[,3]) - 
      plogis(Predict(h2, tpa=1, lp = logxp0)[,3])
l  <- predict(l0, data.frame(lp = logxp0)) - 
      predict(l1, data.frame(lp = logxp0))

plot(x=xp, y=p1exp, type='l', lty=1, lwd=4, xlim=c(0,.35), ylim=c(-0.007,.05), col="red",
     xlab="Baseline risk", ylab="Benefit by tPA", cex.lab=1.2, las=1, bty='l' )
# benefit with interaction terms
lines(x=xp, y=p1, type='l', lty=2, lwd=3, col="dark blue") 
lines(x=xp, y=p2, type='l', lty=3, lwd=2, col="purple") 
lines(x=xp, y=l,  type='l', lty=1, lwd=3, col="black") 

# horizontal line
lines(x=c(0,.5), y=c(0,0)) 
# distribution of predicted 
histSpike(plogis(lp.no.tx), add=T, side=1, nint=300, frac=.1) 

legend("topleft", lty=c(1,2,3,1), pch=c(NA,NA,NA,NA), lwd=c(4,3,2,3), bty='n',
       col=c("red", "dark blue", "purple","black"), cex=1.2,
       legend=c("Expected with proportional effect", 
                "Linear interaction", "Spline smoothing, 2 df",
                "Loess"))

This plot confirms that all estimates of benefit by baseline risk are more or less similar, with benefit clearly increasing by baseline risk. For very low baseline risk, the loess estimates are implausible.

We can also add the grouped observations by decile, as in Califf 1997. The 95% confidence intervals show that the uncertainty per risk group is huge.

Q: How many risk-based groups should be used for illustration of benefit by risk? Default: use quartiles to define 4 quarters; perhaps 3 or only 2 groups in smaller trials?

Code

g <- 10 # first deciles
h  <- lrm(day30 ~ tpa + tpa * lp, data=gusto, eps=0.005, maxit=30)
h2 <- lrm(day30 ~ tpa + rcs(lp,3)*tpa, data=gusto, eps=0.005, maxit=99)
# loess smoothing
l0 <- loess(day30 ~ lp, data=gusto, subset=tpa==0)
l1 <- loess(day30 ~ lp, data=gusto, subset=tpa==1)

# subtract predicted risks with from without tx
p1 <- plogis(Predict(h,  tpa=0, lp = logxp0)[,3]) - 
      plogis(Predict(h,  tpa=1, lp = logxp0)[,3])
p2 <- plogis(Predict(h2, tpa=0, lp = logxp0)[,3]) - 
      plogis(Predict(h2, tpa=1, lp = logxp0)[,3])
l  <- predict(l0, data.frame(lp = logxp0)) - 
      predict(l1, data.frame(lp = logxp0))

plot(x=xp, y=p1exp, type='l', lty=1, lwd=4, xlim=c(0,.35), ylim=c(-0.007,.05), col="red",
     xlab="Baseline risk", ylab="Benefit by tPA", cex.lab=1.2, las=1, bty='l' )
# benefit with interaction terms
lines(x=xp, y=p1, type='l', lty=2, lwd=3, col="dark blue") 
lines(x=xp, y=p2, type='l', lty=3, lwd=2, col="purple") 
lines(x=xp, y=l,  type='l', lty=1, lwd=3, col="black") 

# horizontal line
lines(x=c(0,.5), y=c(0,0)) 
# distribution of predicted 
histSpike(plogis(lp.no.tx), add=T, side=1, nint=300, frac=.1) 

# risk groups
groups <- cut2(lp.no.tx, g=g)
group0 <- groups[gusto$tpa==0]  # SK gropup
group1 <- groups[gusto$tpa==1]  # tPA group

rate0 <- prop.table(table(group0, gusto$day30[gusto$tpa==0]),1 )[,2]
rate1 <- prop.table(table(group1, gusto$day30[gusto$tpa==1]),1 )[,2]
ratediff <- rate0-rate1 # benefit of tPA by group

# CI re-calculation
events1   <- table(group0, gusto$day30[gusto$tpa==0])[,2]
nevents1  <- table(group0, gusto$day30[gusto$tpa==0])[,1]
events2   <- table(group1, gusto$day30[gusto$tpa==1])[,2]
nevents2  <- table(group1, gusto$day30[gusto$tpa==1])[,1]
n1      <- events1 + nevents1
n2      <- events2 + nevents2
CI      <- BinomDiffCI(x1 = events1, n1 = n1, x2 = events2, n2 = n2, method = "scorecc")

points(x=rate0, y=ratediff, pch=1, cex=1, lwd=1, col="black")
arrows(x0=rate0, x1=rate0, y0=CI[,2], y1=CI[,3], angle=90, code=3,len=.1, col="black", lwd=.5)

legend("topleft", lty=c(1,2,3,1,NA), pch=c(NA,NA,NA,NA,1), lwd=c(4,3,2,3,1), bty='n',
       col=c("red", "dark blue", "purple","black","black"), cex=1.2,
       legend=c("Expected with proportional effect", 
                "Linear interaction", "Spline smoothing, 2 df",
                "Loess",
                "Grouped patients"))

Code

g <- 4 # quarters

plot(x=xp, y=p1exp, type='l', lty=1, lwd=4, xlim=c(0,.35), ylim=c(-0.007,.05), col="red",
     xlab="Baseline risk", ylab="Benefit by tPA", cex.lab=1.2, las=1, bty='l' )
# benefit with interaction terms
lines(x=xp, y=p1, type='l', lty=2, lwd=3, col="dark blue") 
lines(x=xp, y=p2, type='l', lty=3, lwd=2, col="purple") 
lines(x=xp, y=l,  type='l', lty=1, lwd=3, col="black") 

# horizontal line
lines(x=c(0,.5), y=c(0,0)) 
# distribution of predicted 
histSpike(plogis(lp.no.tx), add=T, side=1, nint=300, frac=.1) 

# risk groups
groups <- cut2(lp.no.tx, g=g)
group0 <- groups[gusto$tpa==0]  # SK gropup
group1 <- groups[gusto$tpa==1]  # tPA group

rate0 <- prop.table(table(group0, gusto$day30[gusto$tpa==0]),1 )[,2]
rate1 <- prop.table(table(group1, gusto$day30[gusto$tpa==1]),1 )[,2]
ratediff <- rate0-rate1 # benefit of tPA by group

# CI re-calculation
events1   <- table(group0, gusto$day30[gusto$tpa==0])[,2]
nevents1  <- table(group0, gusto$day30[gusto$tpa==0])[,1]
events2   <- table(group1, gusto$day30[gusto$tpa==1])[,2]
nevents2  <- table(group1, gusto$day30[gusto$tpa==1])[,1]
n1      <- events1 + nevents1
n2      <- events2 + nevents2
CI      <- BinomDiffCI(x1 = events1, n1 = n1, x2 = events2, n2 = n2, method = "scorecc")

points(x=rate0, y=ratediff, pch=1, cex=1, lwd=1, col="black")
arrows(x0=rate0, x1=rate0, y0=CI[,2], y1=CI[,3], angle=90, code=3,len=.1, col="black", lwd=.5)

legend("topleft", lty=c(1,2,3,1,NA), pch=c(NA,NA,NA,NA,1), lwd=c(4,3,2,3,1), bty='n',
       col=c("red", "dark blue", "purple","black","black"), cex=1.2,
       legend=c("Expected with proportional effect", 
                "Linear interaction", "Spline smoothing, 2 df",
                "Loess",
                "Grouped patients"))

Conclusions

The GUSTO-I trial serves well to illustrate the impact of conditioning on baseline covariates when we consider relative and absolute effects of treatment on binary outcomes. The risk-adjusted estimate of the overall treatment effect has a different interpretation than the unadjusted estimate: the effect for ‘Patients with acute MI’ versus ‘A patient with a certain risk profile’.

Implications for reporting of RCTs

RCTs typically report on:

an overall effect in the primary analysis;
this analysis should condition on baseline covariates as argued in another blog.
effects stratified by single characteristics: one at a time subgroup analyses;
these analyses should be regarded as secondary and exploratory.

Future RCT reports should include:

an adjusted estimate of the overall treatment effect as the primary analysis;
effects stratified by baseline risk;
typically in 4 risk-based subgroups for illustration,
and in an analysis with continuous baseline risk, typically plotted as benefit by baseline risk.
traditional subgroup analyses only as secondary and exploratory information; not to influence decision-making based on the current RCT, but to inform future studies and new RCTs. An exception may be the situation that strong prior hypotheses exist on effect modeification on the relative scale, as discussed in the ICEMAN report.

References

GUSTO-I references

Gusto Investigators - New England Journal of Medicine, 1993
An international randomized trial comparing four thrombolytic strategies for acute myocardial infarction

Califf R, …, ML Simoons, EJ Topol, GUSTO-I Investigators - American heart journal, 1997
Selection of thrombolytic therapy for individual patients: development of a clinical model

EW Steyerberg, PMM Bossuyt, KL Lee - American heart journal, 2000
Clinical trials in acute myocardial infarction: should we adjust for baseline characteristics?

PATH Statement references

The Predictive Approaches to Treatment effect Heterogeneity (PATH) Statement
David M. Kent, MD, MS; Jessica K. Paulus, ScD; David van Klaveren, PhD; Ralph D’Agostino, PhD; Steve Goodman, MD, MHS, PhD; Rodney Hayward, MD; John P.A. Ioannidis, MD, DSc; Bray Patrick-Lake, MFS; Sally Morton, PhD; Michael Pencina, PhD; Gowri Raman, MBBS, MS; Joseph S. Ross, MD, MHS; Harry P. Selker, MD, MSPH; Ravi Varadhan, PhD; Andrew Vickers, PhD; John B. Wong, MD; and Ewout W. Steyerberg, PhD
Ann Intern Med. 2020;172:35-45.

Annals of Internal Medicine, main text

Annals of Internal Medicine, Explanation and Elaboration

Editorial by Localio et al, 2020

Risks of traditional subgroup analyses

SF Assmann, SJ Pocock, LE Enos, LE Kasten - The Lancet, 2000 Subgroup analysis and other (mis) uses of baseline data in clinical trials

PM Rothwell - The Lancet, 2005 Subgroup analysis in randomised controlled trials: importance, indications, and interpretation

RA Hayward, DM Kent, S Vijan… - BMC medical research nmethodology, 2006 Multivariable risk prediction can greatly enhance the statistical power of clinical trial subgroup analysis

AV Hernández, E Boersma, GD Murray… - American heart journal, 2006 Subgroup analyses in therapeutic cardiovascular clinical trials: are most of them misleading?

JD Wallach, …, KL Sainani, EW Steyerberg, JPA Ioannidis - JAMA internal medicine, 2017 Evaluation of evidence of statistical support and corroboration of subgroup claims in randomized clinical trials

JD Wallach, …, JF Trepanowski, EW Steyerberg, JPA Ioannidis - BMJ, 2016 Sex based subgroup differences in randomized controlled trials: empirical evidence from Cochrane meta-analyses

Reuse

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