Robust Bayesian FDR Control Using Bayes Factors, with Applications to Multi-tissue eQTL Discovery


Motivated by the genomic application of expression quantitative trait loci (eQTL) mapping, we propose a new procedure to perform simultaneous testing of multiple hypotheses using Bayes factors as input test statistics. One of the most significant features of this method is its robustness in controlling the targeted false discovery rate even under misspecifications of parametric alternative models. Moreover, the proposed procedure is highly computationally efficient, which is ideal for treating both complex system and big data in genomic applications. We discuss the theoretical properties of the new procedure and demonstrate its power and computational efficiency in applications of single-tissue and multi-tissue eQTL mapping.

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We thank Debashis Ghosh, Matthew Stephens, and Timothee Flutre for their fruitful discussion and feedbacks. We are grateful for the insightful comments from the two anonymous reviewers. This work is supported by the NIH Grant R01 MH101825 (PI: M.Stephens).

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Correspondence to Xiaoquan Wen.


Appendix 1: Proof of Proposition 1


Suppose that the true generative distributions of \({{\varvec{Y}}}_i\) under the null and alternative model are given by \(p^{0^*}_i\) and \(p^{1^*}_i\), respectively. As sample size \(n_i\) is sufficiently large for each test, the true Bayes factor, \(\mathrm{BF}_i^*\), has the following properties

$$\begin{aligned} \begin{array}{ll} &{}\mathrm{BF}_i^* \xrightarrow {P} 0,~\text{ if } {{\varvec{Y}}}_i \sim p^{0^*}_i, \\ &{}~~~~~~~~~\text {and} \\ &{} \mathrm{BF}_i^* \xrightarrow {P} \infty ,~\text{ if } {{\varvec{Y}}}_i \sim p^{1^*}_i.\\ \end{array} \end{aligned}$$

Assumption 1, in contrast, only requires that the assumed Bayes factors satisfy

$$\begin{aligned} \mathrm{BF}_i \xrightarrow {P} 0,~\text{ if } {{\varvec{Y}}}_i \sim p^{0^*}_i. \end{aligned}$$

Under the conditions stated in Assumption 2, \(\Pr \left( {\hat{\pi }}_0 \ge \pi _0 \mid \pi _0 \right) \rightarrow 1\) implies that \(\Pr \left( {\hat{\pi }}_0 \ge \pi _0 \mid {{\varvec{\mathcal {Y}}}}\right) \rightarrow 1\) for an arbitrary prior distribution on \(\pi _0\). Let \(\epsilon := \Pr \left( {\hat{\pi }}_0 < \pi _0 \mid {{\varvec{\mathcal {Y}}}}\right) \). Then, it follows that

$$\begin{aligned} \epsilon \xrightarrow {P} 0,~\text{ as } \text{ the } \text{ number } \text{ of } \text{ null } \text{ tests } \text{ is } \text{ sufficiently } \text{ large. } \end{aligned}$$

Importantly, the large number of the null tests is required for ensuring the upper-bound property of \({\hat{\pi }}_0\).


$$\begin{aligned} \begin{aligned} \Pr (Z_i = 0 \mid {{\varvec{\mathcal {Y}}}})&= \int \frac{\pi _0}{\pi _0 + (1-\pi _0) \mathrm{BF}_i^*} \, p(\pi _0 \mid {{\varvec{Y}}}_i)\,d\,\pi _0 \\&= \int _{\pi _0 \le {\hat{\pi }}_0} \frac{\pi _0}{\pi _0 + (1-\pi _0) \mathrm{BF}_i^*} \, p(\pi _0 \mid {{\varvec{Y}}}_i)\,d\,\pi _0\\&\quad \,+ \int _{\pi _0 > {\hat{\pi }}_0} \frac{\pi _0}{\pi _0 + (1-\pi _0) \mathrm{BF}_i^*} \, p(\pi _0 \mid {{\varvec{Y}}}_i)\,d\,\pi _0 \\&\le \frac{{\hat{\pi }}_0}{{\hat{\pi }}_0 + (1- {\hat{\pi }}_0) \mathrm{BF}_i^*} + \epsilon \end{aligned} \end{aligned}$$

By (6.1), as \(n_i \rightarrow \infty \),

$$\begin{aligned} \frac{{\hat{\pi }}_0}{{\hat{\pi }}_0 + (1- {\hat{\pi }}_0) \mathrm{BF}_i^*} \xrightarrow {P} I\{{{\varvec{Y}}}_i \sim p^{0*}_i \} + 0 \cdot I\{{{\varvec{Y}}}_i \sim p^{1*}_i \} \end{aligned}$$

whereas by (6.2),

$$\begin{aligned} \begin{array}{ll} &{}\dfrac{{\hat{\pi }}_0}{{\hat{\pi }}_0 + (1- {\hat{\pi }}_0) \mathrm{BF}_i} \xrightarrow {P} 1, ~ \text{ if } ~ {{\varvec{Y}}}_i \sim p^{0*}_i, \\ &{}\dfrac{{\hat{\pi }}_0}{{\hat{\pi }}_0 + (1- {\hat{\pi }}_0) \mathrm{BF}_i} \ge 0, ~ \text{ if } ~ {{\varvec{Y}}}_i \sim p^{1*}_i. \end{array} \end{aligned}$$


$$\begin{aligned} \lim _{n \rightarrow \infty } \Pr \left( \frac{{\hat{\pi }}_0}{{\hat{\pi }}_0 + (1- {\hat{\pi }}_0) \mathrm{BF}_i} \ge \frac{{\hat{\pi }}_0}{{\hat{\pi }}_0 + (1- {\hat{\pi }}_0) \mathrm{BF}_i^*} \right) = 1, \end{aligned}$$

and by (6.3) and (6.4), each individual test satisfies

$$\begin{aligned} \lim _{n_i \rightarrow \infty } \Pr \left( (1 - {\hat{v}}_i) \ge \Pr (Z_i = 0 \mid {{\varvec{Y}}}_i) \right) = 1 \end{aligned}$$

The decision rule stated in the Proposition 1 yields a true Bayesian FDR

$$\begin{aligned} \overline{\text {FDR}} = \frac{\sum _{i=1}^m \delta ^\dagger _i \Pr (Z_i =0 \mid {{\varvec{Y}}}_i)}{D^\dagger \vee 1}. \end{aligned}$$

By (6.8), it is clear that

$$\begin{aligned} \lim _{n \rightarrow \infty } \Pr \left( \overline{\text {FDR}}\le \frac{\sum _{i=1}^m \delta ^\dagger _i (1-{\hat{v}}_i)}{D^\dagger \vee 1} \right) = 1. \end{aligned}$$

Therefore, the Bayesian FDR can be consistently controlled. Furthermore, controlling \(\frac{\sum _{i=1}^m \delta ^\dagger _i (1-{\hat{v}}_i)}{D^\dagger \vee 1} \le \alpha \) ensures that

$$\begin{aligned} \text {FDR}= \mathrm{E}(\overline{\text {FDR}}) \le \mathrm{E}\left( \frac{\sum _{i=1}^m \delta ^\dagger _i (1-{\hat{v}}_i)}{D^\dagger \vee 1} \right) \le \alpha , \end{aligned}$$

because \(\overline{\text {FDR}}\) is obviously bounded. \(\square \)

Appendix 2: Proof of Lemma 2


Let \((\mathrm{BF}_{(1)},\dots ,\mathrm{BF}_{(m)})\) denote the order statistics from the m Bayes factors that are all generated from the respective null hypotheses. Let \(M_j := \frac{1}{j} \sum _{i=1}^j \mathrm{BF}_{(i)}\) denote the partial sample mean computed by the EBF procedure. Note that the sequence \(M_1, M_2, \dots \) is monotonically non-decreasing. Furthermore, by the law of large numbers and the result of Lemma 1, it follows that

$$\begin{aligned} M_m \xrightarrow {P} 1, \end{aligned}$$

for sufficiently large m.

In the case that \(d_0 < m\), it must be true that

$$\begin{aligned} 1 \le M_{d_0+1} \le M_m \xrightarrow {P} 1. \end{aligned}$$

Because the truncated sample mean from the order statistics \((\mathrm{BF}_{(d_0+1)}, ... , \mathrm{BF}_{(m)})\) converges to a quantity that is strictly >1, their contribution to the overall sample mean, \(M_m\), must be negligible, i.e.,

$$\begin{aligned} \frac{m - d_0}{m} \, \left( \frac{1}{m-d_0}\sum _{j=d_0+1}^m \mathrm{BF}_{(j)} \right) \xrightarrow {P} 0. \end{aligned}$$

Taken together, we have shown that

$$\begin{aligned} {\hat{\pi }}_0 \, = \, \frac{d_0}{m} \, \xrightarrow {P} \, 1. \end{aligned}$$

\(\square \)

Appendix 3: Proof of Proposition 2


In the general mixture case, let \(\mathcal {S}^0\) denote the subset of Bayes factors whose data are generated from the null models. Based on Lemma 2, applying the EBF procedure on \(\mathcal {S}^0\) results in an estimate

$$\begin{aligned} \frac{d_{\mathcal {S}^0}}{ |\mathcal {S}^0|} \xrightarrow {P} 1 , \end{aligned}$$

where \(|\mathcal {S}^0|\) denotes the cardinality of \(\mathcal {S}^0\). In the mixed samples, (6.14) suggests that

$$\begin{aligned} M_{d_{\mathcal {S}^0}} = \frac{1}{d_{\mathcal {S}^0}} \, \sum _{j=1}^{d_{\mathcal {S}^0}} \mathrm{BF}_{(j)} \le 1. \end{aligned}$$

The LHS should be strictly <1 if there exists small values of Bayes factors from the alternative models.

Because the EBF procedure finds the largest subset whose sample mean is <1, it must hold true that

$$\begin{aligned} \Pr ( d_0 \ge d_{\mathcal {S}^0} \mid \pi _0 ) \rightarrow 1, \end{aligned}$$

and thus we conclude that

$$\begin{aligned} \Pr ({\hat{\pi }}_0 \ge \pi _0 \mid \pi _0 ) \rightarrow 1. \end{aligned}$$

\(\square \)

Appendix 4: Proof of Corollary 1


Given a predefined FDR level \(\alpha \), the rejection threshold on the estimated false discovery probabilities is given by

$$\begin{aligned} t^\dagger _{\alpha } = \arg \min _t \left( \frac{\sum _{i=1}^m \delta _i^\dagger (t) (1-{\hat{v}}_i )}{D^\dagger (t) \vee 1} \le \alpha \right) . \end{aligned}$$

Equivalently when there is at least one rejection, the above rejection threshold implies that the rejection set \(\Omega := \{i: {\hat{v}}_i > t^\dagger _{\alpha }\}\) is the largest set such that

$$\begin{aligned} \frac{\sum _{ i \in \Omega } \Pr (Z_i =0 \mid {{\varvec{Y}}}_i, {\hat{\pi }}_0)}{|| \Omega || } \le \alpha , \end{aligned}$$

where \(\Pr (Z_i = 0 \mid {{\varvec{Y}}}_i, {\hat{\pi }}_0) = (1-{\hat{v}}_i)\) and \(|| \Omega || = D^\dagger (t_{\alpha }^\dagger )\) denotes the cardinality of the set \(\Omega \). That is, the average estimated false rejection probability in the rejection set should be \(\le \alpha \). Consequently, it implies that if the i-th test yields \(\Pr (Z_i = 0 \mid {{\varvec{Y}}}_i, {\hat{\pi }}_0) \le \alpha \), it must be included in the rejection set (because \(\Omega \) is the largest such set). Therefore, to prove the corollary, it is sufficient to show that \(\Pr (Z_i=0 \mid {{\varvec{Y}}}_i, {\hat{\pi }}_0) \le \alpha \), provided that \(\mathrm{BF}_i \ge \frac{m}{\alpha }\).

Applying the EBF procedure, a single Bayes factor with the value exceeding m leads to

$$\begin{aligned} {\hat{\pi }}_0 \le 1- \frac{1}{m}. \end{aligned}$$

This implies

$$\begin{aligned} \Pr (Z_i= & {} 0 \mid {{\varvec{Y}}}_i, {\hat{\pi }}_0) \le \frac{ 1- \frac{1}{m} }{(1- \frac{1}{m}) + \frac{1}{m} \mathrm{BF}_i} \le \frac{m-1}{m+(m-1) \alpha }\nonumber \\&\cdot \alpha< \frac{m-1}{ m -1 + m \alpha } \cdot \alpha < \alpha \end{aligned}$$

\(\square \)

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Wen, X. Robust Bayesian FDR Control Using Bayes Factors, with Applications to Multi-tissue eQTL Discovery. Stat Biosci 9, 28–49 (2017).

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  • False Discovery Rate
  • Null Model
  • eQTL Mapping
  • False Discovery Rate Control
  • False Discovery Rate Level