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Empirical Priors and Coverage of Posterior Credible Sets in a Sparse Normal Mean Model


Bayesian methods provide a natural means for uncertainty quantification, that is, credible sets can be easily obtained from the posterior distribution. But is this uncertainty quantification valid in the sense that the posterior credible sets attain the nominal frequentist coverage probability? This paper investigates the frequentist validity of posterior uncertainty quantification based on a class of empirical priors in the sparse normal mean model. In particular, we show that our marginal posterior credible intervals achieve the nominal frequentist coverage probability under conditions slightly weaker than needed for selection consistency and a Bernstein–von Mises theorem for the full posterior, and numerical investigations suggest that our empirical Bayes method has superior frequentist coverage probability properties compared to other fully Bayes methods.

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  1. In fact, it is not uncommon in seminar talks or less formal discussions to hear one motivate the construction of a new prior by saying that existing priors “don’t work” and/or the new prior “works better,” not that it more accurately reflects subjective prior beliefs, etc.

  2. The expression for πn(S) given in Section 4.1 of Martin et al. (2017) has a typo, but the correct formula is given in the supplement at


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The authors thank the editors of the special issue of Sankhya A dedicated to Jayanta K. Ghosh for the invitation to contribute, and the anonymous reviewers for their helpful suggestions that improved both our results and presentation. This work is partially supported by the National Science Foundation, DMS–1737933.

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Correspondence to Ryan Martin.

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Appendix : Proof of Theorem 2

Appendix : Proof of Theorem 2

The proof strategy here closely follows that of Theorem 5 in the supplement to Martin et al. (2017) presented in the most recent arXiv version (arXiv:1406.7718). To fix notation, let S be the true configuration of size s = |S|, and let \(S^{\dagger } \subseteq S^{\star }\) be the set of all i such that \(|\theta _{i}^{\star }| \geq \rho _{n}\), where ρn is as in Eq. 2.6, and write s = |S|.

Based on Theorem 2 in Martin et al. (2017), we can restrict to configurations S such that |S| ≤ Cs, where s = |S| and C is a large constant. Take such an S that also satisfies \(S \not \supseteq S^{\dagger }\). Then πn(S) can be bounded as follows:

$$ \pi^{n}(S) \leq \frac{\pi^{n}(S)}{\pi^{n}(S^{\dagger})} = \frac{\pi(S)}{\pi(S^{\dagger})} z^{|S|-s^{\dagger}} e^{\frac{\alpha}{2\sigma^{2}}\{\|Y_{S^{\dagger c}}\|^{2} - \|Y_{S^{c}}\|^{2}\}}, $$

where z = (1 + ατ− 1)− 1/2 < 1. A key observation is that

$$ \|Y_{S^{\dagger c}}\|^{2} - \|Y_{S^{c}}\|^{2} = \|Y_{S \cap S^{\dagger c}}\|^{2} - \|Y_{S^{c} \cap S^{\dagger}}\|^{2}, $$

and the latter two terms are independent since they depend on disjoint sets of Yi’s. Therefore, using this independence and the familiar central and non-central chi-square moment generating functions, we get

$$ \mathsf{E}_{\theta^{\star}} e^{\frac{\alpha}{2\sigma^{2}}\{\|Y_{S^{\dagger c}}\|^{2} - \|Y_{S^{c}}\|^{2}\}} = (1-\alpha)^{-|S \cap S^{\dagger c}|} (1+\alpha)^{-|S^{c} \cap S^{\dagger}|} e^{-\frac{\alpha}{2(1+\alpha)\sigma^{2}} \|\theta_{S^{c} \cap S^{\dagger}}^{\star}\|^{2}}. $$

By definition of S, and the fact that 1 + α > 1, the above expectation can be upper-bounded by

$$ (1-\alpha)^{-|S \cap S^{\dagger c}|} (n^{M})^{-|S^{c} \cap S^{\dagger}|}. $$

Putting the pieces together we have

$$ \mathsf{E}_{\theta^{\star}} \pi^{n}(S) \leq \frac{\pi(S)}{\pi(S^{\dagger})} z^{|S|-s^{\dagger}} (1-\alpha)^{-|S \cap S^{\dagger c}|} (n^{M})^{-|S^{c} \cap S^{\dagger}|}. $$

We want to sum this over all \(S \not \supseteq S^{\dagger }\) but, since it only involves size of S, we only need to sum over sizes. Indeed, after plugging in the definition of π(S) we get

$$ \sum\limits_{S: S \not\supseteq S^{\dagger}, |S| \leq C s^{\star}} \mathsf{E}_{\theta^{\star}} \pi^{n}(S) \leq \sum\limits_{s=0}^{Cs^{\star}} \sum\limits_{t=0}^{s \wedge s^{\dagger}} \frac{\binom{s}{t} \binom{n-s^{\dagger}}{s-t} \binom{n}{s^{\dagger}}}{\binom{n}{s}} \frac{f_{n}(s)}{f_{n}(s^{\dagger})} z^{s-s^{\dagger}} (1-\alpha)^{-(s-t)} (n^{M})^{-(s^{\dagger} - t)}. $$

For the binomial coefficient ratio we have the following simplification and bound:

$$ \frac{\binom{s}{t} \binom{n-s^{\dagger}}{s-t} \binom{n}{s^{\dagger}}}{\binom{n}{s}} = \binom{s}{t} \binom{n-s}{s^{\dagger} - t} \leq s^{s-t} n^{s^{\dagger} - t}. $$

Next, to bound the double-sum, split it into two parts:

$$ \sum\limits_{s=0}^{Cs^{\star}} \sum\limits_{t=0}^{s \wedge s^{\dagger}} (\cdots) = \sum\limits_{s=0}^{s^{\dagger} - 1} \sum\limits_{t=0}^{s} (\cdots) + \sum\limits_{s=s^{\dagger}}^{Cs^{\star}} \sum\limits_{t=0}^{s^{\dagger}} (\cdots) $$

We need to show that both parts on the right-hand side above vanish as \(n \to \infty \). For the first double-sum we have

$$ \sum\limits_{s=0}^{s^{\dagger} - 1} \sum\limits_{t=0}^{s} (\cdots) = \sum\limits_{s=0}^{s^{\dagger} - 1} \left(\frac{1}{K_{1} n^{M-a_{1}-1}} \right)^{s^{\dagger} - s} \sum\limits_{t=0}^{s} \left(\frac{s}{(1-\alpha)n^{M-1}} \right)^{s-t}. $$

Since M > 1 + a1, the inner sum is O(1) and the outer sum—because there is a common n−(Ma1− 1) factor—is o(1) as \(n \to \infty \). Similarly, for the second double-sum we have

$$ \sum\limits_{s=s^{\dagger}}^{Cs^{\star}} \sum\limits_{t=0}^{s^{\dagger}} (\cdots) = \sum\limits_{s=s^{\dagger}}^{Cs^{\star}} \left(\frac{K_{2} s}{(1-\alpha)n^{a_{2}}} \right)^{s-s^{\dagger}} \sum\limits_{t=0}^{s^{\dagger}} \left(\frac{s}{1-\alpha} \right)^{s^{\dagger} - t} \left(\frac{1}{n^{M-1}} \right)^{s^{\dagger} - t}. $$

The inner sum is O(1) and, since a2 < a1, the outer sum is upper-bounded by O(sna2) which goes to 0 by assumption. Both terms in the double-sum above vanish as \(n \to \infty \), thus proving the claim.

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Martin, R., Ning, B. Empirical Priors and Coverage of Posterior Credible Sets in a Sparse Normal Mean Model. Sankhya A 82, 477–498 (2020).

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Keywords and phrases.

  • Bayesian inference
  • Bernstein–von Mises theorem
  • Concentration rate
  • High-dimensional model
  • Uncertainty quantification

AMS (2000) subject classification.

  • Primary 62C12
  • 62F12
  • Secondary 62E20