2.1 Method-of-moments estimators

Moments estimators for meta-analysis are semiparametric; they involve specifying only the first two moments of the distribution of population effects, namely $\mu $ and $\tau ^2$ . Because these methods do not require specifying the higher moments, they do not requiring assuming that population effects are normal. Specifically, consider k studies whose population effects, $\mu _i$ , have expectation $\mu $ and variance $\tau ^2$ . These two moments are the usual meta-analytic estimands of interest. Let $\widehat {\theta }_i$ and $\sigma _i$ respectively denote the point estimate and standard error of the ith study, such that $\widehat {\theta }_i \sim N \left ( \mu _i, \sigma _i^2 \right )$ holds approximately. The within-study standard errors $\sigma _i$ are generally treated as fixed and known.

For a given estimate of the heterogeneity variance, $\widehat {\tau }^2$ , the estimated marginal variance of $\widehat {\theta }_i$ is $\widehat {\tau }^2 + \sigma _i^2$ . The uniformly minimum variance unbiased estimator (UMVUE) of $\mu $ arises from weighting studies by the inverse of their estimated marginal variances,Reference Harville ⁶ denoted $w_i = 1 / (\widehat {\tau }^2 + \sigma _i^2)$ :

$$ \begin{align*} \widehat{\mu} &= \frac{\sum_{i=1}^k w_i \widehat{\theta}_i}{\sum_{i=1}^k w_i}. \end{align*} $$

The various moments estimators are distinguished by their estimators for $\tau ^2$ , and hence the form of the weights $w_i$ . Detailed reviewsReference Langan, Higgins and Jackson ^{7 ,} Reference Veroniki, Jackson, Viechtbauer, Bender, Bowden, Knapp and Salanti ^{36 ,} Reference Viechtbauer ³⁷ and original papers on these approaches are available, so here we summarize briefly. Moments estimators for $\tau ^2$ are based on the generalized Q-statistic:

(1) $$ \begin{align} Q &= \sum_{i=1}^k a_i (\widehat{\theta}_i - \widehat{\mu})^2, \end{align} $$

where the form of the coefficients, $a_i$ , differs across moments estimators. For example, the traditional Dersimonian–Laird estimator (DL)Reference DerSimonian and Laird ¹ sets $a_i = 1 / \sigma _i^2$ . The two-step DL estimator (DL2)Reference DerSimonian and Kacker ² instead sets $a_i = 1 / (\widehat {\tau }^2_{DL} + \sigma _i^2)$ , where $\widehat {\tau }^2_{DL}$ is an initial estimate obtained using the DL estimator. The Paule–Mandel (PM)Reference Paule and Mandel ^{3 ,} Reference van Aert and Jackson ⁴ estimator can be viewed as a limiting case of DL2, involving iteration over the estimates $\widehat {\mu }$ and $\widehat {\tau }^2$ until convergence. This estimator is also equivalent to the empirical Bayes estimator.Reference Raudenbush and Bryk ⁵ In general terms, empirical Bayes estimation uses the observed data to estimate the parameters of the Bayesian prior, rather than specifying the prior independently of the data.Reference Gelman, Carlin and Stern ²¹ In the context of meta-analysis, the empirical Bayes estimator essentially estimates the distribution of population effects by their posterior means, with the prior determined empirically.Reference Raudenbush and Bryk ⁵

2.2 Likelihood-based estimators

In contrast to moments estimators, commonly used likelihood-based estimators assume that the population effects, $\mu _i$ , arise independently from the distribution $\mu _i \sim N \left ( \mu , \tau ^2 \right )$ . Thus, the marginal distribution of studies’ point estimates, $\widehat {\theta }_i$ , is $\widehat {\theta }_i \sim N \left ( \mu , \tau ^2 + \sigma _i^2 \right )$ . We denote the k-vector of point estimates as $\widehat {\boldsymbol {\theta }}$ . Letting $S_i(\tau ) = \sqrt {\tau ^2 + \sigma _i^2}$ be the true marginal standard deviation of the ith study, the joint likelihood is:

(2)

The standard ML estimator for $\tau $ is obtained as usual by solving $\frac {\partial }{\partial \tau } \; \log p ( \widehat {\boldsymbol {\theta }} \mid \mu , \tau ) = 0$ , whose solution depends on $\mu $ .Reference Harville ⁶ Since this estimator does not take into account the loss in degrees of freedom due to the additional estimation of $\mu $ itself, the resulting estimate is often negatively biased.Reference Harville ⁶ This issue motivates REML estimation, which can improve upon ML estimation by transforming the log-likelihood to remove the parameter $\mu $ .Reference Harville ⁶

2.3 Interval estimation

A simple Wald confidence interval can be obtained by assuming $\widehat {\mu }$ is normally distributed, which holds asymptotically in k by standard ML properties. If the weights $w_i$ are treated as known rather than estimated, we have $\widehat {\text {Var}}(\widehat {\mu }) = 1 / \sum _{i=1}^k w_i$ . A Wald 95% confidence interval is:

$$ \begin{align*} \widehat{\mu} &\pm c \sqrt{\widehat{\text{Var}}(\widehat{\mu})}, \end{align*} $$

where $c = \Phi ^{-1}(0.975) \approx 1.96$ is the critical value of the standard normal distribution. However, Wald intervals exhibit substantial under-coverage for small meta-analyses, both because the normal approximation holds only asymptotically and because the approximation $\widehat {\text {Var}}(\widehat {\mu }) = 1 / \sum _{i=1}^k w_i$ does not account for the estimation of $\tau ^2$ .Reference Langan, Higgins and Jackson ^{7 ,} Reference Langan, Higgins and Simmonds ^{11 ,} Reference IntHout, Ioannidis and Borm ¹² Wald intervals can also be constructed for $\widehat {\tau }$ , but exhibit similarly poor performance.Reference Viechtbauer ¹⁹ We therefore do not further discuss Wald intervals, focusing instead on the better-performing alternatives discussed below.

Regarding interval estimation for $\mu $ , the alternative HKSJ, sometimes called “Knapp–Hartung,” interval addresses the limitations of the Wald interval.Reference Sidik and Jonkman ^{13 ,} Reference Knapp and Hartung ¹⁴ This method more flexibly assumes that $\widehat {\mu }$ follows a t distribution and additionally rescales $\widehat {\text {Var}}(\widehat {\mu })$ to account for the estimation of $\tau ^2$ in the weights $w_i$ :

$$ \begin{align*} \widehat{\text{Var}}(\widehat{\mu}) = \frac{\sum_{i=1}^k w_i (\widehat{\theta}_i - \widehat{\mu})^2 }{(k-1) \sum_{i=1}^k w_i }. \end{align*} $$

For $\tau $ , improved intervals can be constructed using the chi-square distribution of the Q statistic, per Eq. (1).Reference Viechtbauer ¹⁹ These “Q-profile” intervals substantially outperform Wald intervals.Reference Viechtbauer ¹⁹ For both $\mu $ and $\tau $ , ML profile intervals can also be constructed in the usual way.Reference Harville ⁶

An interesting, relatively new approach provides exact rather than asymptotic intervals and is theoretically guaranteed to provide more than nominal coverage, under the assumption of normal population effects.Reference Michael, Thornton and Xie ¹⁸ This method essentially involves inverting exact tests. Other parametric methods provide finite-sample corrections to the likelihood ratio test statistic; these include Skovgaard’s second-order correction and Bartlett’s correction.Reference Huizenga, Visser and Dolan ^{38 –} Reference Guolo and Varin ⁴⁰ These methods can improve upon basic likelihood methods for hypothesis testing,Reference Guolo and Varin ⁴⁰ but Skovgaard’s second-order correction was not designed for interval estimation and can be numerically unstable in this context.Reference Kosmidis, Guolo and Varin ³¹ Interval estimation with Bartlett’s correction is possible,Reference Noma ⁴¹ but is not implemented in existing software (I. Visser, personal communication, 8 July 2024).Reference Hilde and Ingmar ^{42 ,} Reference Veroniki, Jackson and Bender ⁴³ Because our focus is on interval estimation rather than testing, our simulations do not include Skovgaard’s or Bartlett’s corrections. Finally, various parametric or nonparametric resampling methods can be used to obtain bootstrapped confidence intervals.Reference Viechtbauer ^{19 ,} Reference Veroniki, Jackson and Bender ^{43 ,} Reference Van Den Noortgate and Onghena ⁴⁴ Nonparametric resampling can be conducted by resampling rows with replacement, after which one can obtain simple percentile bootstrap intervals or bias-corrected and accelerated (BCa) intervals, among many other types of bootstrap intervals.Reference Efron ^{45 ,} Reference Carpenter and Bithell ⁴⁶ The BCa confidence corrects for bias and skewness in the bootstrapped sampling distribution, which we speculate could be helpful when estimating the sampling distribution of $\tau $ . The BCa bootstrap has performed relatively well for certain meta-analytic estimators that are functions of $\widehat {\tau }$ .Reference Mathur and VanderWeele ⁴⁷ However, bootstrapping is an asymptotic procedure whose finite-sample performance typically must be assessed through simulations.

2.4 Existing simulations comparing these methods

Langan et al. (2017)Reference Langan, Higgins and Simmonds ¹¹ provide an excellent systematic review of simulation studies for different heterogeneity estimators.Reference Langan, Higgins and Jackson ⁷ Briefly, the DL estimator was negatively biased for $\tau $ when heterogeneity was moderate to high, and the PM estimator was typically less biased.Reference Langan, Higgins and Simmonds ¹¹ The reviewed studies do not appear to have assessed interval estimation for $\tau $ . Based on their own, more extensive simulation study, Langan et al. (2019)Reference Langan, Higgins and Jackson ⁷ generally recommend REML, PM, or DL2 for heterogeneity estimation, along with HKSJ confidence intervals for $\mu $ ; however, they recommend caution in interpreting heterogeneity estimates in small meta-analyses.

Langan et al.’s (2019)Reference Langan, Higgins and Jackson ⁷ simulation study did not assess intervals based on the profile likelihood, bootstrapping, or the exact method; the latter was developed only recently. Regarding profile intervals, recommendations in the literature are inconsistent. A prominent paper stated that “the profile likelihood is a good method for computing confidence intervals”.Reference Cornell, Mulrow and Localio ⁴⁸ One simulation study seemed to support this recommendation, finding that when the heterogeneity is greater than zero, profile likelihood intervals showed the closest to nominal coverage.Reference Kontopantelis and Reeves ¹⁰ On the other hand, another simulation study suggested that profile intervals often exhibited under-coverage for meta-analyses of only 5 studies.Reference Guolo ³⁹ The originators of the exact method provide simulations suggesting that the resulting intervals are not substantially wider than those of existing methods, despite the method’s theoretical guarantee of at least nominal coverage.Reference Michael, Thornton and Xie ¹⁸ While our simulation study is primarily motivated by investigating the Jeffreys methods, a secondary contribution is to more extensively evaluate profile, bootstrap, and exact intervals. We now turn to establishing the theory for the Jeffreys1 and Jeffreys2 priors.

3.1 The Jeffreys priors

Under the assumption of normal population effects, Bodnar et al. (2017)Reference Bodnar, Link and Arendacká ¹⁵ showed that the improper Jeffreys1 prior is:

$$ \begin{align*} p \left( \mu, \tau \right) \propto \tau \sqrt{\sum_{i=1}^k \left(S_i(\tau)\right)^{-4}} \end{align*} $$

where, again, $S_i(\tau ) = \sqrt {\tau ^2 + \sigma _i^2}$ . Since this prior is independent of $\mu $ , it can be expressed as two independent priors on $\mu $ and $\tau $ , where the prior on $\mu $ is uniform:

(3) $$ \begin{align} p \left( \mu, \tau \right) &\propto p(\mu) \; p(\tau), \text{ where } p(\mu) \propto 1 \text{ and } p(\tau) \propto \tau \sqrt{\sum_{i=1}^k \left(S_i(\tau)\right)^{-4}}. \end{align} $$

If $\mu $ is treated as the only parameter of interest and $\tau $ is considered a nuisance parameter, then the Jeffreys1 prior also coincides with the Berger–Bernardo reference prior.Reference Bodnar, Link and Elster ³⁰ In general, the Berger–Bernardo prior for a given distribution is designed to be maximally “noninformative” in the sense of minimizing the amount of information provided by the prior and maximizing the amount of information provided by the data.Reference Bodnar, Link and Elster ^{30 ,} Reference Berger and Bernardo ⁴⁹ Specifically, this prior maximizes the Kullback–Liebler divergence between the prior and the posterior.Reference Berger and Bernardo ⁴⁹

Regarding the Jeffreys2 prior, the joint likelihood in Eq. (2) implies that the entries of the expected Fisher information are:

$$ \begin{align*} k_{\mu\mu} := E \Bigg[ \frac{\partial^2 \ell}{\partial\mu^2} \Bigg] = -\sum_{i=1}^k \left(S_i(\tau)\right)^{-2}, \; \; \; k_{\tau \tau} := E \Bigg[ \frac{\partial^2 \ell}{\partial \tau^2} \Bigg] = -2 \tau^2 \sum_{i=1}^k \left(S_i(\tau)\right)^{-4}, \; \; \; k_{\mu \tau} := E \Bigg[ \frac{\partial^2 \ell}{\partial\mu\partial \tau} \Bigg] = 0 \end{align*} $$

where $\ell $ is the likelihood function. Therefore, the Jeffreys2 prior is $p \left ( \mu , \tau \right ) \propto \sqrt { k_{\mu \mu } k_{\tau \tau } }$ . (This result is straightforward to show directly, or alternatively can be viewed as a simple special case of the prior given in Mathur (2024).Reference Mathur ^{34 ,} Footnote ⁱⁱ This yields the improper two-parameter prior:

$$ \begin{align*} p \left( \mu, \tau \right) &\propto \tau \sqrt{\left( \sum_{i=1}^k \left(S_i(\tau)\right)^{-2} \right) \left( \sum_{i=1}^k \left(S_i(\tau)\right)^{-4} \right)}. \end{align*} $$

Like the Jeffreys1 prior, the Jeffreys2 prior can be expressed as:

(4) $$ \begin{align} p \left( \mu, \tau \right) &\propto p(\mu) \; p(\tau), \text{ where } p(\mu) \propto 1 \text{ and } p(\tau) \propto \tau \sqrt{\left( \sum_{i=1}^k \left(S_i(\tau)\right)^{-2} \right) \left( \sum_{i=1}^k \left(S_i(\tau)\right)^{-4} \right)}. \end{align} $$

To illustrate, Figure 1 shows both priors on $\tau $ for four meta-analyses of standardized mean differences. The meta-analyses were simulated with studies’ sample sizes, N, arising from four different distributions. Although the magnitude of the priors will of course be affected by the number of studies k, their shape is minimally affected by k, so Figure 1 depicts the prior for meta-analyses with $k=10$ . Note that for each meta-analysis, the Jeffreys2 prior is somewhat narrower than the Jeffreys1 prior, suggesting that the former may provide narrower intervals; this hypothesis will be explored in more depth in the simulation study (Section 4). Both priors lead to proper posteriors if $k> 1$ (see Bodnar (2017)Reference Bodnar, Link and Arendacká ¹⁵ regarding Jeffreys1 and the present Section 1 of the Supplementary Material, regarding Jeffreys2). Additionally, both priors generalize easily to the case of meta-regression: the Jeffreys1 prior would coincide with that of Bodnar et al. (2024) for generalized marginal random effects models,Reference Bodnar and Bodnar ⁵⁰ and we derive the Jeffreys2 prior for meta-regression in Section 1 of the Supplementary Material. We do not further consider meta-regression in the main text.

Figure 1

Priors for four simulated meta-analyses of standardized mean differences ( $k=10$ ), in which the within-study sample sizes (N) were generated from four possible distributions. Studies’ standard errors were estimated using Eq. (5) and, given the data-generation parameters, were approximately equal to $2/\sqrt {N}$ . Points are the maxima. The priors have been scaled to have the same maximum height.

3.2 The posterior under each prior

For either prior, since $p(\mu , \tau ) \propto p(\tau )$ , the marginal posterior on $\tau $ is:Reference Bodnar, Link and Arendacká ¹⁵

In turn, the conditional posterior of $\mu $ , given $\tau $ , is normal:Reference Röver ^{9 ,} Reference Bodnar, Link and Arendacká ^{15 ,} Reference Gelman, Carlin and Stern ²¹

$$ \begin{align*} p(\mu \mid \tau, \widehat{\boldsymbol{\theta}}) &\propto \frac{1}{\sqrt{2\pi \; \text{Var}(\mu \mid \tau, \widehat{\boldsymbol{\theta}})} } \cdot \exp \Bigg\{ -\frac{1}{2 \; \text{Var}(\mu \mid \tau, \widehat{\boldsymbol{\theta}})} \left( \mu - E[ \mu \mid \tau, \widehat{\boldsymbol{\theta}}] \right)^2 \Bigg\}, \end{align*} $$

where:

$$ \begin{align*} E[ \mu \mid \tau, \widehat{\boldsymbol{\theta}}] = \frac{\sum_{i=1}^k \widehat{\theta}_i \left(S_i(\tau)\right)^{-2} }{\sum_{i=1}^k \left(S_i(\tau)\right)^{-2}}, \; \; \text{Var}(\mu \mid \tau, \widehat{\boldsymbol{\theta}}) &= \frac{1}{\sum_{i=1}^k \left(S_i(\tau)\right)^{-2}}. \end{align*} $$

Thus, the joint posterior $p(\mu , \tau \mid \widehat {\boldsymbol {\theta }})$ can be decomposed into the two tractable components $p(\tau \mid \widehat {\boldsymbol {\theta }})$ and $p(\mu \mid \tau , \widehat {\boldsymbol {\theta }})$ .Reference Röver ⁹ Given this observation, Röver and othersReference Röver ^{9 ,} Reference Röver and Friede ⁵¹ developed theory and software for a discrete approximation to the joint posterior $p(\mu , \tau \mid \widehat {\boldsymbol {\theta }})$ and the marginal posterior on $\mu $ , given by the mixture distribution:

$$ \begin{align*} p(\mu \mid \widehat{\boldsymbol{\theta}}) &\propto \int_{0}^{\infty} p(\mu \mid \tau, \widehat{\boldsymbol{\theta}}) \; p(\tau \mid \widehat{\boldsymbol{\theta}}) \; d\tau. \end{align*} $$

The discrete approximation approach does not require sampling via mixed-chain Monte Carlo (MCMC) and is implemented in the R package bayesmeta.Reference Röver ^{9 ,} Reference Röver and Friede ⁵¹ We use this package in our simulations and applied example.

With approximations to the joint and marginal posteriors in hand, point estimates can be defined in terms of various measures of central tendency, such as the posterior mode, median, or mean. For either prior, $p(\mu \mid \widehat {\boldsymbol {\theta }})$ appears to be nearly symmetric in many cases (e.g., Figure 4), so the three measures of central tendency will often agree closely. However, this is not the case for $p(\tau \mid \widehat {\boldsymbol {\theta }})$ , which is asymmetric under either prior. Existing work on the Jeffreys1 prior focused primarily on posterior means and medians,Reference Bodnar, Link and Arendacká ¹⁵ but we focus on posterior modes given their aforementioned theoretical advantages.Reference Firth ²⁶ Indeed, as discussed in Section 4.4, our simulations indicated that posterior modes for $\tau $ provided substantially lower bias, root mean square error (RMSE), and mean absolute error (MAE) than did posterior means and medians. As in ML estimation, point estimates can be defined either in terms of the marginal or the joint mode. In the Bayesian context, the marginal mode represents the value of a given parameter (e.g., $\mu $ ) that maximizes the posterior for that parameter alone, marginalizing over the other parameter (e.g., $\tau $ ). In contrast, the joint mode represents the values of both parameters that jointly maximize the joint posterior. We consider marginal modes in this paper to provide a more direct comparison to marginal ML estimation, which is the usual implementation for meta-analysis.

Figure 2

Priors on $\tau $ for the meta-analysis on all-cause death ( $k=3, \{ \sigma _i \} = \{ 1.15, 1.63, 0.19 \}$ ). Points are maxima. The priors have been scaled to have the same maximum height.

Figure 3

Joint posterior under the Jeffreys2 prior for the meta-analysis on all-cause death ( $k=3, \{ \sigma _i \} = \{ 1.15, 1.63, 0.19 \}$ ). Horizontal red line: marginal posterior mode of $\mu $ . Vertical blue line: marginal posterior mode of $\tau $ .

Figure 4

Marginal posteriors under the Jeffreys2 prior for the meta-analysis on all-cause death. Solid vertical lines: marginal posterior modes. Dashed vertical lines: limits of 95% intervals.

Figure 5

Interval limits greater than $RR=10$ are truncated. The exact method does not yield point estimates. CI: credible interval.

Also analogously to ML estimation, symmetric Wald credible intervals are sometimes constructed for Bayesian estimates by approximating the posterior as asymptotically normal around the posterior mode, with a variance–covariance matrix equal to the inverse of the Hessian of the negative log-posterior evaluated at the posterior mode.Reference Gelman, Carlin and Stern ²¹ However, just as Wald intervals around the ML estimate can perform poorly if the likelihood is asymmetric, Wald intervals around the posterior mode can likewise perform poorly if the posterior is asymmetric.Reference Greenland ⁵² To obtain appropriately asymmetric posterior intervals, we consider two approaches. First, a central (also called “equal-tailed”) 95% posterior quantile interval can be obtained by taking the 2.5th and 97.5th quantiles of the estimated posterior distribution. Second, the shortest possible 95% posterior quantile interval can be obtained numerically; this interval is equivalent to a highest posterior density interval for unimodal distributions.Reference Gelman, Carlin and Stern ²¹ In our simulations and applied example, we obtain both types of intervals from the R package bayesmeta.Reference Röver ⁹

3.3 Theoretical and substantive distinctions between the priors

The distinction between the Jeffreys1 and Jeffreys2 priors invokes theoretical and substantive considerations that pertain in general to multiparameter Jeffreys priors. Jeffreys and others have argued that multiparameter Jeffreys priors are appropriate if one wishes to estimate all of the parameters (i.e., both $\mu $ and $\tau $ in meta-analysis), but not if one wishes to estimate only a subset of the parameters (i.e., only $\mu $ ), with the others treated as nuisance parameters.Reference Kass and Wasserman ^{24 ,} Reference Bernardo ^{25 ,} Reference Jeffreys ⁵³ As noted in the Introduction, a random-effects meta-analysis should generally involve estimation and reporting of $\tau $ (or related metricsReference Schünemann, Higgins, Vist, Higgins, Thomas, Chandler, Cumpston, Li, Page and Welch ^{32 ,} Reference Riley, Higgins and Deeks ^{54 ,} Reference Mathur and VanderWeele ⁵⁵ ) in addition to $\mu $ , which suggests consideration of the Jeffreys2 prior. On the other hand, in general location-scale problems, Jeffreys recommended obtaining the prior with respect to only the scale parameters, holding constant the location parameters.Reference Kass and Wasserman ^{24 ,} Reference Jeffreys ⁵³ This would correspond to the Jeffreys1 prior. Jeffreys’ recommendation was motivated by problems that can arise when the number of location parameters increases with the sample size, similarly to the well-known Neyman–Scott problem in which the ML estimator fails to be consistent.Reference Kass and Wasserman ^{24 ,} Reference Jeffreys ⁵³ Interestingly, Firth later showed that in a specific, severe version of the Neyman–Scott problem, the multiparameter Jeffreys prior (i.e., the Firth correction) in fact leads to a consistent and exactly unbiased estimator.Reference Firth ²⁶ This was unexpected given that the asymptotic arguments justifying the Firth correction are violated with an increasing number of parameters.Reference Firth ²⁶ Of course, in the present setting of random-effects meta-analysis, the number of parameters is fixed, so this potential issue does not arise in the first place. Our view is that existing substantive and theoretical considerations do not clearly rule out either prior as inappropriate for random-effects meta-analysis, so our simulation study evaluates both.

4.1 Point and interval estimation methods

Table 1 lists the methods assessed in our simulation study. We assessed both Jeffreys priors. For point estimation under each prior, we primarily considered marginal posterior modes but secondarily investigated posterior means and medians (Section 2.2 of the Supplementary Material). Regarding interval estimation for $\mu $ , central and shortest intervals were generally quite similar, so we only show results for shortest intervals. Regarding interval estimation for $\tau $ , we consider both types of intervals for each prior, termed “Jeffreys1-shortest,” “Jeffreys1-central,” “Jeffreys2-shortest,” and “Jeffreys2-central.”

Table 1

Methods assessed in simulation study

$^{a}$ In pilot tests for the scenarios with $k=10$ , the bootstrap methods were not competitive with other methods, so these computationally intensive methods were not run for other sample sizes.

We compared the performance of both Jeffreys priors to that of several existing frequentist methods that were described in Section 2. We selected methods that have performed well in existing, large simulation studies or that have desirable theoretical properties, such as providing appropriately asymmetric intervals for $\tau $ .Reference Harville ^{6 ,} Reference Langan, Higgins and Jackson ^{7 ,} Reference Michael, Thornton and Xie ^{18 ,} Reference Guolo ^{39 ,} Reference Cornell, Mulrow and Localio ^{48 ,} Reference Brockwell and Gordon ⁵⁷ For point estimation, the comparison methods were ML estimation, REML, DL, DL2, and PM. Regarding interval estimation for $\mu $ , we considered HKSJ intervals for each frequentist estimation method, ML profile intervals (ML-profile), exact intervals,Reference Michael, Thornton and Xie ¹⁸ nonparametric BCa bootstrap intervals, and nonparametric percentile bootstrap intervals.Reference Efron ^{45 ,} Reference Carpenter and Bithell ⁴⁶ Regarding interval estimation for $\tau $ , we considered Q-profile intervals for each frequentist estimation method, as well as ML-profile and both bootstrap intervals. We implemented all frequentist methods and intervals using the R package metafor Reference Viechtbauer ⁵⁸ with the following exceptions: we implemented ML-profile using custom R code, the exact method using the R package rma.exact,Reference Michael, Thornton and Xie ¹⁸ and the bootstrap methods using the R package boot.Reference Canty and Boot ⁵⁹

4.2 Data generation

Table 2 summarizes the simulation parameters we manipulated, which were similar to those of Langan et al.’s (2019) simulation study.Reference Langan, Higgins and Jackson ⁷ We considered continuous outcomes with point estimates on the Hedges’ g scaleReference Hedges ⁵⁶ as well as binary outcomes with point estimates on the log-odds ratio scale. We considered both normally distributed and exponentially distributed population effects; in the latter case, the assumptions for all point estimators except the moments estimators were violated. Statistical theory suggests that all methods would perform comparably for very large meta-analyses with normal effects, and accordingly our focus is on point and interval estimation for smaller meta-analyses ( $k \le 20$ ). Our primary simulations reported in the main text are those with $k \in \{ 2, 3, 5, 10, 20\}$ . We additionally ran simulations with $k=100$ to confirm asymptotic behavior (Section 3 of the Supplementary Material). Because the bootstrap intervals required much more computational time than the other methods, we first pilot-tested them in all scenarios with a single sample size ( $k=10$ ) to assess whether these methods were competitive with other methods.

Table 2

Possible values of simulation parameters

$^{a}$ Results for scenarios with $k=100$ appear in the Supplementary Material; these scenarios are excluded from aggregated results in the main text.

Data generation proceeded as follows. For each simulation iterate, we generated a meta-analysis whose underlying population effects ( $\mu _i$ ) were either normal or exponential. Normal population effects were generated as $\mu _i \sim N \left ( \mu , \tau ^2 \right )$ , where we varied $\mu $ and $\tau $ as indicated in Table 2. Exponential population effects were generated from an appropriately scaled and shifted distribution to achieve the desired population moments, $\mu $ and $\tau ^2$ . For each study in the meta-analysis, we generated a total sample size N from one of the four distributions listed in Table 2. We then simulated individual participant data, such that $N/2$ participants were allocated to a treatment group, and the other $N/2$ to a control group. In scenarios with a continuous outcome, we simulated outcomes with a mean of 0 in the control group and $\mu _i$ in the treatment group, and with a standard deviation of 1 within each group. We then estimated the standardized mean difference using the Hedges’ g correction.Reference Hedges ^{56 ,} Reference Viechtbauer ⁵⁸ We used the standard large-sample approximation for studies’ standard errors (Eq. (8) in Hedges (1982)Reference Hedges ⁶⁰ ):

(5) $$ \begin{align} \widehat{\sigma}_i = \sqrt{ \frac{N^c + N^t}{N^c N^t} + \frac{\widehat{\theta}_i^2}{2(N^c + N^t)} } = \sqrt{ \frac{8 + \widehat{\theta}_i^2}{2N} } \end{align} $$

where $N^c$ and $N^t$ are the within-group sample sizes, which were both equal to $N/2$ in our simulations. The expectation of this estimator is approximately $\sqrt {(8 + \mu ^2)/2N}$ .

In scenarios with a binary outcome, we simulated outcomes from a logistic model such that:

$$ \begin{align*} P(Y = 1 \mid X) = \text{expit} \Big\{ \text{logit}\{ P(Y = 1 \mid X = 0) \} + \mu_i X \Big\} \end{align*} $$

where $P(Y = 1 \mid X = 0)$ was a scenario parameter that we manipulated among the values listed in Table 2. We then estimated the odds ratio; to handle potential zero cell counts when present, we added 0.5 to each table cell when any cells had a count of zero.Reference Viechtbauer ⁵⁸

We expected that for binary outcomes and small within-study sample sizes, certain extreme combinations of scenario parameters (e.g., $N=40$ and $\mu =2.3$ , corresponding to an extreme odds ratio of 10) would result in biased within-study odds ratios.Reference Firth ^{26 ,} Reference Nemes, Jonasson and Genell ⁶¹ In pilot simulations, we identified combinations of scenario parameters that resulted in within-study absolute bias of greater than 0.05. We excluded these combinations of scenario parameters since our focus is on bias arising from meta-analytic estimation methods rather than from within-study bias. After excluding these combinations of simulation parameters, we ultimately simulated 240 unique scenarios for continuous outcomes and 2267 for binary outcomes.

4.3 Performance metrics

For each scenario, we assessed the point estimators’ performance and variability in terms of their bias, MAE, and RMSE, defined in the usual frequentist sense. That is, for a generic parameter $\omega _r$ that varies across 500 simulation iterates, r:

$$ \begin{align*} \text{Bias} &= \frac{1}{500} \sum_{r=1}^{500} \left( \widehat{\omega}_r - \omega_r \right) \\ \text{MAE} &= \frac{1}{500} \sum_{r=1}^{500} \vert \widehat{\omega}_r - \omega_r \vert \\ \text{RMSE} &= \left( \frac{1}{500} \sum_{r=1}^{500} \left( \widehat{\omega}_r - \omega_r \right)^2 \right)^{1/2}. \end{align*} $$

For each scenario, we assessed interval estimation in terms of the frequentist coverage and width of 95% confidence or credible intervals. Some methods’ intervals exhibited over-coverage in some scenarios but exhibited under-coverage in others. Therefore, when aggregating results across scenarios, we also consider the percentage of scenarios in which each method achieved approximately nominal coverage, defined stringently as having coverage $>$ 94%. In the Discussion, we expand upon our reasons for assessing frequentist properties of Bayesian methods, and the implications of this approach. We did not assess statistical power. Although p-values can certainly be useful when interpreted as continuous measures of evidence, we concur with others’ longstanding concerns about bright-line significance testing,Reference Rothman ^{62 ,} Reference McShane, Gal and Gelman ⁶³ a practice that has contributed to striking misinterpretations of published meta-analysesReference Mathur and VanderWeele ^{55 ,} Reference Mathur and VanderWeele ⁶⁴ and likely also to publication bias.

4.5 Overall conclusions

All methods performed similarly for point estimation of $\mu $ . In general, standard frequentist methods with HKSJ intervals for $\mu $ and Q-profile intervals for $\tau $ performed the most consistently across outcome types. Jeffreys2-shortest also showed consistently strong performance for meta-analyses of binary outcomes, and yielded substantially narrower intervals than frequentist methods. However, the Jeffreys2-shortest interval did not perform as consistently for continuous outcomes; this method exhibited mild under-coverage for very small meta-analyses with high heterogeneity. Regarding point estimation of $\tau $ , all methods again performed comparably on average, though the optimal method depended on the value of $\tau $ itself. Regarding interval estimation for $\tau $ , the Q-profile method arguably performed the best and behaved consistently across scenarios.

Overall, for small meta-analyses of continuous outcomes, we would recommend standard frequentist methods with an HKSJ interval for $\mu $ and a Q-profile interval for $\tau $ , consistent with previous recommendations. However, for small meta-analyses of binary outcomes, Jeffreys2 may be preferable over standard frequentist methods if the meta-analyst is primarily interested in point and interval estimation for $\mu $ , potentially along with point estimation of $\tau $ (although, again, the best-performing method for estimating $\tau $ depended on the value of $\tau $ itself). This is because the Jeffreys2-shortest interval more frequently had at least nominal coverage, yet was substantially more precise. If the meta-analyst is also interested in obtaining an interval for $\tau $ , then using a frequentist method with a Q-profile interval for $\tau $ would likely provide closer to nominal coverage for $\tau $ than would Jeffrey2-shortest; however, this would likely sacrifice a substantial amount of precision for $\mu $ .