Abstract
Estimating causal effects from randomized experiments is central to clinical research. Reducing the statistical uncertainty in these analyses is an important objective for statisticians. Registries, prior trials, and health records constitute a growing compendium of historical data on patients under standardofcare that may be exploitable to this end. However, most methods for historical borrowing achieve reductions in variance by sacrificing strict typeI error rate control. Here, we propose a use of historical data that exploits linear covariate adjustment to improve the efficiency of trial analyses without incurring bias. Specifically, we train a prognostic model on the historical data, then estimate the treatment effect using a linear regression while adjusting for the trial subjects’ predicted outcomes (their prognostic scores). We prove that, under certain conditions, this prognostic covariate adjustment procedure attains the minimum variance possible among a large class of estimators. When those conditions are not met, prognostic covariate adjustment is still more efficient than raw covariate adjustment and the gain in efficiency is proportional to a measure of the predictive accuracy of the prognostic model above and beyond the linear relationship with the raw covariates. We demonstrate the approach using simulations and a reanalysis of an Alzheimer’s disease clinical trial and observe meaningful reductions in meansquared error and the estimated variance. Lastly, we provide a simplified formula for asymptotic variance that enables power calculations that account for these gains. Sample size reductions between 10% and 30% are attainable when using prognostic models that explain a clinically realistic percentage of the outcome variance.
Acknowledgments
We are grateful to Xinkun Nie and Oleg Sofrygin for enlightening conversations and to Rachael C. Aikens for feedback on a draft of this article. Data collection and sharing for this project was funded in part by the Alzheimer’s Disease Neuroimaging Initiative (ADNI) (National Institutes of Health Grant U01 AG024904) and DOD ADNI (Department of Defense award number W81XWH1220012). ADNI is funded by the National Institute on Aging, the National Institute of Biomedical Imaging and Bioengineering, and through generous contributions from the following: AbbVie, Alzheimer’s Association; Alzheimer’s Drug Discovery Foundation; Araclon Biotech; BioClinica, Inc.; Biogen; BristolMyers Squibb Company; CereSpir, Inc.; Cogstate; Eisai Inc.; Elan Pharmaceuticals, Inc.; Eli Lilly and Company; EuroImmun; F. HoffmannLa Roche Ltd and its affiliated company Genentech, Inc.; Fujirebio; GE Healthcare; IXICO Ltd.; Janssen Alzheimer Immunotherapy Research & Development, LLC.; Johnson & Johnson Pharmaceutical Research & Development LLC.; Lumosity; Lundbeck; Merck & Co., Inc.; Meso Scale Diagnostics, LLC.; NeuroRx Research; Neurotrack Technologies; Novartis Pharmaceuticals Corporation; Pfizer Inc.; Piramal Imaging; Servier; Takeda Pharmaceutical Company; and Transition Therapeutics. The Canadian Institutes of Health Research is providing funds to support ADNI clinical sites in Canada. Private sector contributions are facilitated by the Foundation for the National Institutes of Health (www.fnih.org). The grantee organization is the Northern California Institute for Research and Education, and the study is coordinated by the Alzheimer’s Therapeutic Research Institute at the University of Southern California. ADNI data are disseminated by the Laboratory for Neuro Imaging at the University of Southern California. Data collection and sharing for this project was funded in part by the University of California, San Diego Alzheimer’s Disease Cooperative Study (ADCS) (National Institute on Aging Grant Number U19AG010483).

Author contribution: All the authors have accepted responsibility for the entire content of this submitted manuscript and approved submission.

Research funding: None declared.

Conflict of interest statement: The authors declare no conflicts of interest regarding this article.
Appendix A. Mathematical results
Throughout we assume enough regularity conditions for the asymptotic normality of Mestimators to hold. The details are found in chapter 5 (thm 5.23) of van der Vaart [49].
Lemma A.1
(Rosenblum). The influence function for the linear regression treatment effect estimator we describe in Section 3 is ψ = ψ_{1} − ψ_{0} where
and
This follows from results in Robins et al. [50]. An accessible presentation for the case of generalized linear models is given in Rosenblum and Laan [51].
Definition A.1
(Differenceinmeans). The “differenceinmeans” (or “unadjusted”) estimator of τ = μ_{1} − μ_{0} is
Note that throughout the appendix we omit the subscript n on estimators. E.g. τ_{Δ} is shorthand for τ_{Δ,n} and our asymptotic statements refer to the sequence of estimators as n becomes large.
Lemma A.2
The differenceinmeans estimator has asymptotic variance given by
where
Proof
This fact is wellknown. One proof follows the outline of 7 below taking Z^{ ⊤ } = [1, W]. □
Definition A.2
(ANCOVA I). The “ANCOVA I” estimator of τ = μ_{1} − μ_{0} (denoted
Definition A.3
(ANCOVA II). The “ANCOVA II” estimator of τ = μ_{1} − μ_{0} (denoted
The following two Theorems A.3 and A.4 are mild generalizations of or follow closely from results stated in Leon et al. [24] and Yang and Tsiatis [16]. Details are provided here for the reader’s convenience.
Theorem A.3
The ANCOVA I estimator is asymptotically unbiased for τ = μ_{1} − μ_{0} and has asymptotic variance given by
where
Proof
We begin by applying Lemma A.1. Minimization of the expected loglikelihood shows that
where
Where
It is known that all regular and asymptotically linear estimators of the treatment effect have an influence function of this form with h(X) dependent on the choice of estimator [24, 26].
By the theory of influence functions, our estimator has a limiting distribution [26]
The asymptotic variance of
The covariance of the two terms involves the expectations
where we have introduced ξ_{*} = π_{1}ξ_{0} + π_{0}ξ_{1}. Assembling obtains the desired result. □
Corollary A.3.1
When X ∈ R (a single covariate), a consistent estimate of the sampling variance
where
Proof
This follows from the definitions and Slutsky’s theorem. □
Corollary A.3.2
If either π_{0} = π_{1} or ξ_{0} = ξ_{1}, then
Theorem A.4
The ANCOVA II estimator is asymptotically unbiased for τ = μ_{1} − μ_{0} and has asymptotic variance given by
Proof
Arguments similar to those in Theorem A.3 show that the influence function for the GLM marginal effect estimator with this specification is identical to Eq. (12) except that ξ = π_{0}ξ_{0} + π_{1}ξ_{1} is replaced by ξ_{*} = π_{1}ξ_{0} + π_{0}ξ_{1}. Specifically ψ_{II} = ψ_{1,II} − ψ_{0,II} with
The result follows from proceeding along the outline of Theorem A.3. □
Corollary A.4.1
When X ∈ R (a single covariate), a consistent estimate of the sampling variance
Corollary A.4.2
Adding covariates to the ANCOVA II estimator can only decrease its asymptotic variance.
Proof
Consider using covariates X with variance Σ_{
x
} and covariance with Y_{
w
} of ξ_{w,x} versus a set of covariates [X, M] (
The denominator must be positive because
Theorem A.5
ANCOVA II is a more efficient estimator than ANCOVA I or differenceinmeans. ANCOVA I may or may not be more efficient than differenceinmeans (unless π_{0} = π_{1} = 0.5 or ξ_{0} = ξ_{1}, in which case it is as efficient as ANCOVA II). In a slight abuse of notation,
Proof
Lemma A.6
Consider using the ANCOVA II estimator with an arbitrary (multivariate) transformation of the covariates f(X) in place of the raw covariates X. Among all fixed transformations f(X), the transformation
Consider replacing X in the interacted linear model (ANCOVA II) with an arbitrary fixed (possibly multivariate) function of the covariates f(X). By Eq. (23) and our definitions of ξ_{*} and V the influence function for this estimator is ψ = ψ_{1} − ψ_{0} with
where
The result is precisely the efficient influence function for the treatment effect [24, 26]. It is known that no regular and asymptotically linear (RAL) estimator (which essentially all practical and reasonable estimators are) can be more efficient than any estimator with this influence function.
Corollary A.6.1
Presume a constant treatment effect: μ_{1}(X) = μ_{0}(X) + τ. Then the ANCOVA II analysis that uses μ_{0}(X) in the role of X has the lowest possible asymptotic variance among all regular and asymptotically linear estimators with access to the covariates X.
Proof
μ_{1}(X) = μ_{0}(X) + τ implies
which is the same as the efficient influence function when μ_{1}(X) = μ_{0}(X) + τ. □
Corollary A.6.2
Corollary A.6.1 also holds when the ANCOVA II estimator is replaced by the ANCOVA I estimator.
Proof
Theorem A.5 establishes that ANCOVA I is as efficient as ANCOVA II when
The following lemma is required for the proof that proceeds it.
Lemma A.7
Let
Proof
The final convergence holds by our assumption that
Taking advantage of the fact that f, f_{
n
} ≤ b are bounded we can make similar arguments to show that
Corollary A.7.1
Let
Proof
Let
Now note
as desired. □
Theorem A.8
Presume X has compact support and there is a constant treatment effect: μ_{1}(X) = μ_{0}(X) + τ with μ_{0}(x) < b bounded. Let m(x) be a (random) function learned from the external data (
Y
′,
X
′)_{n′} such that m(x) < b is also bounded and
Proof
Define our estimator of interest as the ANCOVA II estimator that uses the learned model m(X) in place of the covariates X if m(X) is not numerically constant up to some machine precision and otherwise as the differenceinmeans estimator. Denote this estimator
Showing
where
Let
To wit, consider the difference
where we’ve abbreviated
And show it converges to 0. Recalling that m itself is random (depends on the external data (
X
′
Y
′)), but independent of the trial data (
X
,
W
,
Y
), note that we can treat m(⋅) as if it were a fixed function and B as a fixed constant if we condition on the external data. After conditioning, the quantity inside the parentheses is IID and has mean zero because its μ_{0}(X) − m(X)B and
where we’ve used the fact that the summands are IID to pass the variance through the sum and effectively gain the 1/n required to cancel the n. The same argument shows that the equivalent for the second term in Eq. (35) is
To complete the proof we invoke Corollary A.7.1 in combination with our assumptions m(x) < b, μ_{0}(x) < b and
Corollary A.8.1
Theorem A.8 also holds for the ANCOVA I estimator.
Proof
In the case of a constant treatment effect ANCOVA I and ANCOVA II have the same asymptotic variance (Theorem A.5). The result follows immediately. □
Appendix B. Estimating
σ
w
2
and ρ_{
w
} for power calculations
One method for obtaining estimates for the marginal potential outcome variances (
The controlarm marginal outcome variance
The correlation ρ_{0} between M″ and Y″ can be estimated by
which is the usual sample correlation coefficient. These values may be inflated (
The corresponding values for the treatment arm can rarely be estimated from data because treatmentarm data for the experimental treatment is likely to be scarce or unavailable. It is therefore prudent to assume
Appendix C. Additional simulation results
Here we detail a full set of simulation results using additional specifications for the regression estimators (Figure 1). “Covariates” indicates whether the raw covariates were adjusted for. “Prognostic score” indicates whether any prognostic score was used, and, if so, whether it was estimated from a training dataset or whether the true value was used. “Interactions” specifies whether treatment × (covariates and/or prognostic score) interactions were used. “SE” indicates the standard deviation of the mean squared error.
Scenario  Covariates  Prognostic score  Interaction  MSE  SE 

Baseline  False  None  True  7.64 × 10^{−2}  1.08 × 10^{−3} 
Baseline  False  None  False  7.64 × 10^{−2}  1.08 × 10^{−3} 
Baseline  False  Estimated  True  1.76 × 10^{−2}  2.46 × 10^{−4} 
Baseline  False  Estimated  False  1.75 × 10^{−2}  2.45 × 10^{−4} 
Baseline  False  Oracle  True  7.69 × 10^{−3}  1.09 × 10^{−4} 
Baseline  False  Oracle  False  7.69 × 10^{−3}  1.09 × 10^{−4} 
Baseline  True  None  True  5.07 × 10^{−2}  7.18 × 10^{−4} 
Baseline  True  None  False  5.04 × 10^{−2}  7.14 × 10^{−4} 
Baseline  True  Estimated  True  1.74 × 10^{−2}  2.46 × 10^{−4} 
Baseline  True  Estimated  False  1.73 × 10^{−2}  2.44 × 10^{−4} 
Baseline  True  Oracle  True  7.85 × 10^{−3}  1.11 × 10^{−4} 
Baseline  True  Oracle  False  7.85 × 10^{−3}  1.11 × 10^{−4} 
Surrrogate  False  None  True  7.47 × 10^{−2}  1.05 × 10^{−3} 
Surrrogate  False  None  False  7.47 × 10^{−2}  1.05 × 10^{−3} 
Surrrogate  False  Estimated  True  4.05 × 10^{−2}  5.69 × 10^{−4} 
Surrrogate  False  Estimated  False  4.03 × 10^{−2}  5.66 × 10^{−4} 
Surrrogate  False  Oracle  True  8.25 × 10^{−3}  1.18 × 10^{−4} 
Surrrogate  False  Oracle  False  8.24 × 10^{−3}  1.18 × 10^{−4} 
Surrrogate  True  None  True  5.03 × 10^{−2}  7.09 × 10^{−4} 
Surrrogate  True  None  False  5.00 × 10^{−2}  7.04 × 10^{−4} 
Surrrogate  True  Estimated  True  3.75 × 10^{−2}  5.27 × 10^{−4} 
Surrrogate  True  Estimated  False  3.72 × 10^{−2}  5.23 × 10^{−4} 
Surrrogate  True  Oracle  True  8.41 × 10^{−3}  1.20 × 10^{−4} 
Surrrogate  True  Oracle  False  8.41 × 10^{−3}  1.20 × 10^{−4} 
Shifted  False  None  True  7.65 × 10^{−2}  1.10 × 10^{−3} 
Shifted  False  None  False  7.65 × 10^{−2}  1.10 × 10^{−3} 
Shifted  False  Estimated  True  6.79 × 10^{−2}  9.62 × 10^{−4} 
Shifted  False  Estimated  False  6.79 × 10^{−2}  9.62 × 10^{−4} 
Shifted  False  Oracle  True  8.20 × 10^{−3}  1.15 × 10^{−4} 
Shifted  False  Oracle  False  8.20 × 10^{−3}  1.15 × 10^{−4} 
Shifted  True  None  True  5.03 × 10^{−2}  7.11 × 10^{−4} 
Shifted  True  None  False  5.00 × 10^{−2}  7.05 × 10^{−4} 
Shifted  True  Estimated  True  4.91 × 10^{−2}  6.97 × 10^{−4} 
Shifted  True  Estimated  False  4.86 × 10^{−2}  6.90 × 10^{−4} 
Shifted  True  Oracle  True  8.34 × 10^{−3}  1.17 × 10^{−4} 
Shifted  True  Oracle  False  8.34 × 10^{−3}  1.17 × 10^{−4} 
Strong  False  None  True  7.73 × 10^{−2}  1.08 × 10^{−3} 
Strong  False  None  False  7.73 × 10^{−2}  1.08 × 10^{−3} 
Strong  False  Estimated  True  1.85 × 10^{−2}  2.65 × 10^{−4} 
Strong  False  Estimated  False  1.85 × 10^{−2}  2.64 × 10^{−4} 
Strong  False  Oracle  True  8.16 × 10^{−3}  1.16 × 10^{−4} 
Strong  False  Oracle  False  8.16 × 10^{−3}  1.16 × 10^{−4} 
Strong  True  None  True  5.14 × 10^{−2}  7.18 × 10^{−4} 
Strong  True  None  False  5.11 × 10^{−2}  7.13 × 10^{−4} 
Strong  True  Estimated  True  1.84 × 10^{−2}  2.62 × 10^{−4} 
Strong  True  Estimated  False  1.82 × 10^{−2}  2.59 × 10^{−4} 
Strong  True  Oracle  True  8.33 × 10^{−3}  1.18 × 10^{−4} 
Strong  True  Oracle  False  8.32 × 10^{−3}  1.18 × 10^{−4} 
Linear  False  None  True  3.49 × 10^{−2}  4.83 × 10^{−4} 
Linear  False  None  False  3.49 × 10^{−2}  4.83 × 10^{−4} 
Linear  False  Estimated  True  9.64 × 10^{−3}  1.38 × 10^{−4} 
Linear  False  Estimated  False  9.64 × 10^{−3}  1.38 × 10^{−4} 
Linear  False  Oracle  True  8.20 × 10^{−3}  1.16 × 10^{−4} 
Linear  False  Oracle  False  8.20 × 10^{−3}  1.16 × 10^{−4} 
Linear  True  None  True  8.37 × 10^{−3}  1.18 × 10^{−4} 
Linear  True  None  False  8.37 × 10^{−3}  1.18 × 10^{−4} 
Linear  True  Estimated  True  8.39 × 10^{−3}  1.19 × 10^{−4} 
Linear  True  Estimated  False  8.39 × 10^{−3}  1.19 × 10^{−4} 
Linear  True  Oracle  True  8.37 × 10^{−3}  1.18 × 10^{−4} 
Linear  True  Oracle  False  8.37 × 10^{−3}  1.18 × 10^{−4} 
Heterogeneous  False  None  True  5.54 × 10^{−2}  7.76 × 10^{−4} 
Heterogeneous  False  None  False  5.54 × 10^{−2}  7.76 × 10^{−4} 
Heterogeneous  False  Estimated  True  2.30 × 10^{−2}  3.23 × 10^{−4} 
Heterogeneous  False  Estimated  False  2.32 × 10^{−2}  3.25 × 10^{−4} 
Heterogeneous  False  Oracle  True  2.29 × 10^{−2}  3.20 × 10^{−4} 
Heterogeneous  False  Oracle  False  2.32 × 10^{−2}  3.24 × 10^{−4} 
Heterogeneous  True  None  True  2.99 × 10^{−2}  4.30 × 10^{−4} 
Heterogeneous  True  None  False  2.98 × 10^{−2}  4.29 × 10^{−4} 
Heterogeneous  True  Estimated  True  2.13 × 10^{−2}  3.01 × 10^{−4} 
Heterogeneous  True  Estimated  False  2.19 × 10^{−2}  3.08 × 10^{−4} 
Heterogeneous  True  Oracle  True  1.89 × 10^{−2}  2.69 × 10^{−4} 
Heterogeneous  True  Oracle  False  1.98 × 10^{−2}  2.81 × 10^{−4} 
Figure 1:
Appendix D. Covariates in the empirical demonstration dataset
Table 4:
Covariate  Description 

AChEI or memantine usage  Whether a subject is using a class of symptomatic Alzheimer’s drugs 
ADAS commands  Assesses the subject’s ability to follow commands 
ADAS comprehension  Assesses the subject’s ability to understand spoken language 
ADAS construction  Assesses the subject’s ability to draw basic figures 
ADAS ideational  Assesses the subject’s ability to carry out a basic task 
ADAS naming  Assesses the subject’s ability to name common objects 
ADAS orientation  Assesses the subject’s knowledge of time and place 
ADAS remember instructions  Assesses the subject’s ability to remember test instructions 
ADAS spoken language  Assesses the subject’s ability to speak clearly 
ADAS word finding  Assesses the subject’s word finding in speech 
ADAS word recall  Assesses the subject’s ability to recall a list of words 
ADAS word recognition  Assesses the subject’s ability to remember and identify words 
Age  Subject age at baseline 
ApoE e4 Allele count  The number of ApoE e4 alleles a subject has (0, 1, or 2) 
CDR community  Assesses the subject’s engagement in community activities 
CDR home and hobbies  Assesses the subject’s engagement in home and personal activities 
CDR judgement  Assesses the subject’s judgement skills 
CDR memory  Assesses the subject’s memory 
CDR orientation  Assesses the subject’s knowledge of time and place 
CDR personal care  Assesses the subject’s ability to care for themselves 
Diastolic blood pressure  The diastolic blood pressure of a subject 
Education (Years)  The number of years of education of a subject 
Heart rate  The resting heart rate of a subject 
Height  The height of a subject 
Indicator for clinical trial  1 if the subject is in an RCT, 0 if not 
MMSE attention and calculation  Assesses the subject’s attention and calculation skills 
MMSE language  Assesses the subject’s language skills 
MMSE orientation  Assesses the subject’s knowledge of place and time 
MMSE recall  Assesses the subject’s ability to remember prompts 
MMSE registration  Assesses the subject’s ability to repeat prompts 
Region: Europe  1 if the subject lives in Europe, 0 otherwise 
Region: Northern America  1 if the subject lives in the US or Canada, 0 otherwise 
Region: Other  1 if the subject lives outside of Europe/US/Canada, 0 otherwise 
Serious adverse events  The number of serious adverse events reported 
Sex  1 if female, 0 if male 
Systolic blood pressure  The systolic blood pressure of a subject 
Weight  The weight of a subject 
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