Augmented IPW (doubly-robust estimation) with causatr

Code

library(causatr)
library(tinytable)

Augmented inverse probability weighting (AIPW) combines the two single-robust estimators — g-computation (an outcome model) and IPW (a treatment model) — into one doubly-robust estimator. For a binary treatment and intervention value $a$,

\[ \hat\psi_{\mathrm{AIPW}}(a) = \underbrace{\frac{1}{n}\sum_i \hat m(a, L_i)}_{\text{g-computation}} + \underbrace{\frac{1}{n}\sum_i \frac{\mathbb 1\{A_i = a\}}{\hat g(a \mid L_i)}\,\bigl(Y_i - \hat m(A_i, L_i)\bigr)}_{\text{IPW augmentation}}, \]

where $\hat m$ is the fitted outcome regression and $\hat g$ the fitted propensity. The augmentation term has mean zero when either model is correct, so $\hat\psi_{\mathrm{AIPW}}$ is consistent if either the outcome model or the propensity model is correctly specified — not necessarily both. That is the double-robustness property.

causatr fits both nuisances with the user’s model_fn (outcome) and propensity_model_fn (treatment), assembles the doubly-robust functional per intervention, and computes the variance with the stacked influence-function sandwich (outcome block + propensity block + plug-in). This is the classical analytical AIPW — distinct from lmtp’s TMLE/SDR with cross-fitting and machine learning.

Double robustness, demonstrated

The cleanest way to see double robustness is a simulated data generating process with a known average treatment effect, where a confounder $L$ enters both the propensity and the outcome nonlinearly. A model that omits the $L^2$ term is misspecified.

Code

set.seed(2)
n <- 4000
L <- rnorm(n)
A <- rbinom(n, 1, plogis(-0.5 + 0.7 * L + 0.6 * L^2))
Y <- 1 + 2 * A + 1.5 * L + 1.6 * L^2 + rnorm(n)
d <- data.frame(Y, A, L)

correct <- ~ L + I(L^2) # captures the L^2 confounding
wrong <- ~ L # misspecified: omits L^2

The true ATE is 2. causatr lets the outcome and treatment models carry separate confounder formulas (confounders_outcome, confounders_treatment), so we can misspecify one while keeping the other correct:

Code

ate <- function(estimator, co, ct) {
  fit <- causat(
    d, outcome = "Y", treatment = "A",
    confounders_outcome = co, confounders_treatment = ct,
    estimator = estimator, model_fn = stats::glm,
    propensity_model_fn = stats::glm
  )
  contrast(fit, list(a1 = static(1), a0 = static(0)), reference = "a0")$contrasts$estimate[1]
}

results <- data.frame(
  estimator = c(
    "gcomp", "ipw",
    "aipw", "aipw", "aipw", "aipw"
  ),
  outcome_model = c(
    "wrong", "correct",
    "wrong", "correct", "correct", "wrong"
  ),
  propensity_model = c(
    "wrong", "wrong",
    "correct", "wrong", "correct", "wrong"
  ),
  ATE_hat = c(
    ate("gcomp", wrong, wrong),
    ate("ipw", correct, wrong),
    ate("aipw", wrong, correct),
    ate("aipw", correct, wrong),
    ate("aipw", correct, correct),
    ate("aipw", wrong, wrong)
  )
)
tt(results, digits = 3)

estimator	outcome_model	propensity_model	ATE_hat
gcomp	wrong	wrong	3.13
ipw	correct	wrong	3.42
aipw	wrong	correct	2.05
aipw	correct	wrong	1.97
aipw	correct	correct	1.96
aipw	wrong	wrong	3.47

Reading the table: g-computation with a misspecified outcome model is biased, and IPW with a misspecified propensity is biased. But AIPW recovers the true ATE of 2 whenever at least one nuisance is correct — even though the other is wrong. Only when both are misspecified does AIPW lose consistency (last row): double robustness buys one free misspecification, not two.

Real-data example: NHEFS

On observational data we never know the truth, but AIPW is the natural default when you are unsure which nuisance you trust. Using the NHEFS quit-smoking question (effect of qsmk on weight change wt82_71):

Code

data("nhefs")
nhefs_complete <- nhefs[!is.na(nhefs$wt82_71) & !is.na(nhefs$education), ]

conf <- ~ sex + age + I(age^2) + race + factor(education) +
  smokeintensity + I(smokeintensity^2) + smokeyrs + I(smokeyrs^2) +
  factor(exercise) + factor(active) + wt71 + I(wt71^2)

fit_aipw <- causat(
  nhefs_complete,
  outcome = "wt82_71", treatment = "qsmk",
  confounders = conf,
  estimator = "aipw",
  model_fn = stats::glm,
  propensity_model_fn = stats::glm
)

res_aipw <- contrast(
  fit_aipw,
  interventions = list(quit = static(1), continue = static(0)),
  reference = "continue",
  type = "difference",
  ci_method = "sandwich"
)
tt(tidy(res_aipw), digits = 3)

term	estimate	std.error	type	conf.low	conf.high
quit vs continue	3.48	0.483	contrast	2.53	4.42

The point estimate sits between the pure g-computation and pure IPW estimates and shares their interpretation: the average weight change had everyone quit smoking versus had no one quit.

Variance

The "sandwich" CI above is the stacked influence-function variance: the AIPW influence function is the sum of the outcome-model correction, the propensity correction, and the doubly-robust plug-in residual, aggregated by variance_if(). It is asymptotically exact under correct specification of the models being used. ci_method = "bootstrap" refits both nuisances on each resample and is available as a non-parametric alternative.

Code

res_boot <- contrast(
  fit_aipw,
  interventions = list(quit = static(1), continue = static(0)),
  reference = "continue",
  ci_method = "bootstrap", n_boot = 200L
)
tt(res_boot$contrasts[, c("comparison", "estimate", "se", "ci_lower", "ci_upper")], digits = 3)

comparison	estimate	se	ci_lower	ci_upper
quit vs continue	3.48	0.482	2.53	4.42

Where AIPW fits

AIPW is a member of the methodological triangle alongside g-computation and IPW: same estimand, different reliance on the nuisance models. Reach for it when

you want a single estimate that is robust to misspecifying one of the two nuisance models, or
you are triangulating: agreement between gcomp, IPW, and AIPW is reassuring, and AIPW disagreeing with both flags a likely misspecification in one of them.

causatr’s AIPW supports the same surface as the other estimators — binary, continuous, categorical, and multivariate treatments; static / shift / scale_by / dynamic / stochastic interventions; difference / ratio / OR contrasts; by-stratified estimands; stabilized weights; and transportability. For the longitudinal doubly-robust estimator (ICE-AIPW, Bang & Robins 2005), see vignette("longitudinal"). For the triangulation workflow across all three estimators, see vignette("triangulation").

References

Robins JM, Rotnitzky A, Zhao LP (1994). Estimation of regression coefficients when some regressors are not always observed. Journal of the American Statistical Association 89:846–866.

Bang H, Robins JM (2005). Doubly robust estimation in missing data and causal inference models. Biometrics 61:962–973.

Hernán MA, Robins JM (2025). Causal Inference: What If. Chapman & Hall/CRC.

--- title: "Augmented IPW (doubly-robust estimation) with causatr" code-fold: show code-tools: true vignette: > %\VignetteIndexEntry{Augmented IPW (doubly-robust estimation) with causatr} %\VignetteEngine{quarto::html} %\VignetteEncoding{UTF-8} --- ```{r} #| include: false knitr::opts_chunk$set(collapse = TRUE, comment = "#>") ``` ```{r} #| message: false library(causatr) library(tinytable) ``` Augmented inverse probability weighting (AIPW) combines the two single-robust estimators — g-computation (an **outcome** model) and IPW (a **treatment** model) — into one **doubly-robust** estimator. For a binary treatment and intervention value $a$, $$ \hat\psi_{\mathrm{AIPW}}(a) = \underbrace{\frac{1}{n}\sum_i \hat m(a, L_i)}_{\text{g-computation}} + \underbrace{\frac{1}{n}\sum_i \frac{\mathbb 1\{A_i = a\}}{\hat g(a \mid L_i)}\,\bigl(Y_i - \hat m(A_i, L_i)\bigr)}_{\text{IPW augmentation}}, $$ where $\hat m$ is the fitted outcome regression and $\hat g$ the fitted propensity. The augmentation term has mean zero when *either* model is correct, so $\hat\psi_{\mathrm{AIPW}}$ is consistent if **either** the outcome model **or** the propensity model is correctly specified — not necessarily both. That is the **double-robustness** property. causatr fits both nuisances with the user's `model_fn` (outcome) and `propensity_model_fn` (treatment), assembles the doubly-robust functional per intervention, and computes the variance with the stacked influence-function sandwich (outcome block + propensity block + plug-in). This is the *classical analytical* AIPW — distinct from `lmtp`'s TMLE/SDR with cross-fitting and machine learning. ## Double robustness, demonstrated The cleanest way to see double robustness is a simulated data generating process with a **known** average treatment effect, where a confounder $L$ enters both the propensity and the outcome **nonlinearly**. A model that omits the $L^2$ term is misspecified. ```{r} set.seed(2) n <- 4000 L <- rnorm(n) A <- rbinom(n, 1, plogis(-0.5 + 0.7 * L + 0.6 * L^2)) Y <- 1 + 2 * A + 1.5 * L + 1.6 * L^2 + rnorm(n) d <- data.frame(Y, A, L) correct <- ~ L + I(L^2) # captures the L^2 confounding wrong <- ~ L # misspecified: omits L^2 ``` The true ATE is **2**. causatr lets the outcome and treatment models carry *separate* confounder formulas (`confounders_outcome`, `confounders_treatment`), so we can misspecify one while keeping the other correct: ```{r} ate <- function(estimator, co, ct) { fit <- causat( d, outcome = "Y", treatment = "A", confounders_outcome = co, confounders_treatment = ct, estimator = estimator, model_fn = stats::glm, propensity_model_fn = stats::glm ) contrast(fit, list(a1 = static(1), a0 = static(0)), reference = "a0")$contrasts$estimate[1] } results <- data.frame( estimator = c( "gcomp", "ipw", "aipw", "aipw", "aipw", "aipw" ), outcome_model = c( "wrong", "correct", "wrong", "correct", "correct", "wrong" ), propensity_model = c( "wrong", "wrong", "correct", "wrong", "correct", "wrong" ), ATE_hat = c( ate("gcomp", wrong, wrong), ate("ipw", correct, wrong), ate("aipw", wrong, correct), ate("aipw", correct, wrong), ate("aipw", correct, correct), ate("aipw", wrong, wrong) ) ) tt(results, digits = 3) ``` Reading the table: g-computation with a misspecified **outcome** model is biased, and IPW with a misspecified **propensity** is biased. But AIPW recovers the true ATE of 2 whenever **at least one** nuisance is correct — even though the other is wrong. Only when **both** are misspecified does AIPW lose consistency (last row): double robustness buys one free misspecification, not two. ## Real-data example: NHEFS On observational data we never know the truth, but AIPW is the natural default when you are unsure which nuisance you trust. Using the NHEFS quit-smoking question (effect of `qsmk` on weight change `wt82_71`): ```{r} data("nhefs") nhefs_complete <- nhefs[!is.na(nhefs$wt82_71) & !is.na(nhefs$education), ] conf <- ~ sex + age + I(age^2) + race + factor(education) + smokeintensity + I(smokeintensity^2) + smokeyrs + I(smokeyrs^2) + factor(exercise) + factor(active) + wt71 + I(wt71^2) fit_aipw <- causat( nhefs_complete, outcome = "wt82_71", treatment = "qsmk", confounders = conf, estimator = "aipw", model_fn = stats::glm, propensity_model_fn = stats::glm ) res_aipw <- contrast( fit_aipw, interventions = list(quit = static(1), continue = static(0)), reference = "continue", type = "difference", ci_method = "sandwich" ) tt(tidy(res_aipw), digits = 3) ``` The point estimate sits between the pure g-computation and pure IPW estimates and shares their interpretation: the average weight change had everyone quit smoking versus had no one quit. ## Variance The `"sandwich"` CI above is the stacked influence-function variance: the AIPW influence function is the sum of the outcome-model correction, the propensity correction, and the doubly-robust plug-in residual, aggregated by `variance_if()`. It is asymptotically exact under correct specification of the models being used. `ci_method = "bootstrap"` refits both nuisances on each resample and is available as a non-parametric alternative. ```{r} res_boot <- contrast( fit_aipw, interventions = list(quit = static(1), continue = static(0)), reference = "continue", ci_method = "bootstrap", n_boot = 200L ) tt(res_boot$contrasts[, c("comparison", "estimate", "se", "ci_lower", "ci_upper")], digits = 3) ``` ## Where AIPW fits AIPW is a member of the methodological **triangle** alongside g-computation and IPW: same estimand, different reliance on the nuisance models. Reach for it when - you want a single estimate that is robust to misspecifying *one* of the two nuisance models, or - you are triangulating: agreement between gcomp, IPW, and AIPW is reassuring, and AIPW disagreeing with both flags a likely misspecification in one of them. causatr's AIPW supports the same surface as the other estimators — binary, continuous, categorical, and multivariate treatments; `static` / `shift` / `scale_by` / `dynamic` / `stochastic` interventions; difference / ratio / OR contrasts; `by`-stratified estimands; stabilized weights; and transportability. For the **longitudinal** doubly-robust estimator (ICE-AIPW, Bang & Robins 2005), see `vignette("longitudinal")`. For the triangulation workflow across all three estimators, see `vignette("triangulation")`. ## References Robins JM, Rotnitzky A, Zhao LP (1994). Estimation of regression coefficients when some regressors are not always observed. *Journal of the American Statistical Association* 89:846–866. Bang H, Robins JM (2005). Doubly robust estimation in missing data and causal inference models. *Biometrics* 61:962–973. Hernán MA, Robins JM (2025). *Causal Inference: What If*. Chapman & Hall/CRC.