Index: The Book of Statistical Proofs ▷ Statistical Models ▷ Univariate normal data ▷ Bayesian linear regression ▷ Log model evidence

Theorem: Let

\[\label{eq:GLM} m: \; y = X \beta + \varepsilon, \; \varepsilon \sim \mathcal{N}(0, \sigma^2 V)\]

be a linear regression model with measured $n \times 1$ data vector $y$, known $n \times p$ design matrix $X$, known $n \times n$ covariance structure $V$ as well as unknown $p \times 1$ regression coefficients $\beta$ and unknown noise variance $\sigma^2$. Moreover, assume a normal-gamma prior distribution over the model parameters $\beta$ and $\tau = 1/\sigma^2$:

\[\label{eq:GLM-NG-prior} p(\beta,\tau) = \mathcal{N}(\beta; \mu_0, (\tau \Lambda_0)^{-1}) \cdot \mathrm{Gam}(\tau; a_0, b_0) \; .\]

Then, the log model evidence for this model is

\[\label{eq:GLM-NG-LME} \begin{split} \log p(y|m) = \frac{1}{2} & \log |P| - \frac{n}{2} \log (2 \pi) + \frac{1}{2} \log |\Lambda_0| - \frac{1}{2} \log |\Lambda_n| + \\ & \log \Gamma(a_n) - \log \Gamma(a_0) + a_0 \log b_0 - a_n \log b_n \end{split}\]

where the posterior hyperparameters are given by

\[\label{eq:GLM-NG-post-par} \begin{split} \mu_n &= \Lambda_n^{-1} (X^\mathrm{T} P y + \Lambda_0 \mu_0) \\ \Lambda_n &= X^\mathrm{T} P X + \Lambda_0 \\ a_n &= a_0 + \frac{n}{2} \\ b_n &= b_0 + \frac{1}{2} (y^\mathrm{T} P y + \mu_0^\mathrm{T} \Lambda_0 \mu_0 - \mu_n^\mathrm{T} \Lambda_n \mu_n) \; . \end{split}\]

Proof: According to the law of marginal probability, the model evidence for this model is:

\[\label{eq:GLM-NG-ME-s1} p(y|m) = \iint p(y|\beta,\tau) \, p(\beta,\tau) \, \mathrm{d}\beta \, \mathrm{d}\tau \; .\]

According to the law of conditional probability, the integrand is equivalent to the joint likelihood:

\[\label{eq:GLM-NG-ME-s2} p(y|m) = \iint p(y,\beta,\tau) \, \mathrm{d}\beta \, \mathrm{d}\tau \; .\]

Equation \eqref{eq:GLM} implies the following likelihood function

\[\label{eq:GLM-LF-class} p(y|\beta,\sigma^2) = \mathcal{N}(y; X \beta, \sigma^2 V) = \sqrt{\frac{1}{(2 \pi)^n |\sigma^2 V|}} \, \exp\left[ -\frac{1}{2 \sigma^2} (y-X\beta)^\mathrm{T} V^{-1} (y-X\beta) \right]\]

which, for mathematical convenience, can also be parametrized as

\[\label{eq:GLM-LF-Bayes} p(y|\beta,\tau) = \mathcal{N}(y; X \beta, (\tau P)^{-1}) = \sqrt{\frac{|\tau P|}{(2 \pi)^n}} \, \exp\left[ -\frac{\tau}{2} (y-X\beta)^\mathrm{T} P (y-X\beta) \right]\]

using the noise precision $\tau = 1/\sigma^2$ and the $n \times n$ precision matrix $P = V^{-1}$.


When deriving the posterior distribution $p(\beta,\tau|y)$, the joint likelihood $p(y,\beta,\tau)$ is obtained as

\[\label{eq:GLM-NG-LME-s1} \begin{split} p(y,\beta,\tau) = \; & \sqrt{\frac{\tau^n |P|}{(2 \pi)^n}} \, \sqrt{\frac{\tau^p |\Lambda_0|}{(2 \pi)^p}} \, \frac{ {b_0}^{a_0}}{\Gamma(a_0)} \, \tau^{a_0-1} \exp[-b_0 \tau] \cdot \\ & \exp\left[ -\frac{\tau}{2} \left( (\beta-\mu_n)^T \Lambda_n (\beta-\mu_n) + (y^T P y + \mu_0^T \Lambda_0 \mu_0 - \mu_n^T \Lambda_n \mu_n) \right) \right] \; . \end{split}\]

Using the probability density function of the multivariate normal distribution, we can rewrite this as

\[\label{eq:GLM-NG-LME-s2} \begin{split} p(y,\beta,\tau) = \; & \sqrt{\frac{\tau^n |P|}{(2 \pi)^n}} \, \sqrt{\frac{\tau^p |\Lambda_0|}{(2 \pi)^p}} \, \sqrt{\frac{(2 \pi)^p}{\tau^p |\Lambda_n|}} \, \frac{ {b_0}^{a_0}}{\Gamma(a_0)} \, \tau^{a_0-1} \exp[-b_0 \tau] \cdot \\ & \mathcal{N}(\beta; \mu_n, (\tau \Lambda_n)^{-1}) \, \exp\left[ -\frac{\tau}{2} (y^T P y + \mu_0^T \Lambda_0 \mu_0 - \mu_n^T \Lambda_n \mu_n) \right] \; . \end{split}\]

Now, $\beta$ can be integrated out easily:

\[\label{eq:GLM-NG-LME-s3} \begin{split} \int p(y,\beta,\tau) \, \mathrm{d}\beta = \; & \sqrt{\frac{\tau^n |P|}{(2 \pi)^n}} \, \sqrt{\frac{|\Lambda_0|}{|\Lambda_n|}} \, \frac{ {b_0}^{a_0}}{\Gamma(a_0)} \, \tau^{a_0-1} \exp[-b_0 \tau] \cdot \\ & \exp\left[ -\frac{\tau}{2} (y^T P y + \mu_0^T \Lambda_0 \mu_0 - \mu_n^T \Lambda_n \mu_n) \right] \; . \end{split}\]

Using the probability density function of the gamma distribution, we can rewrite this as

\[\label{eq:GLM-NG-LME-s4} \int p(y,\beta,\tau) \, \mathrm{d}\beta = \sqrt{\frac{|P|}{(2 \pi)^n}} \, \sqrt{\frac{|\Lambda_0|}{|\Lambda_n|}} \, \frac{ {b_0}^{a_0}}{\Gamma(a_0)} \, \frac{\Gamma(a_n)}{ {b_n}^{a_n}} \, \mathrm{Gam}(\tau; a_n, b_n) \; .\]

Finally, $\tau$ can also be integrated out:

\[\label{eq:GLM-NG-LME-s5} \iint p(y,\beta,\tau) \, \mathrm{d}\beta \, \mathrm{d}\tau = \sqrt{\frac{|P|}{(2 \pi)^n}} \, \sqrt{\frac{|\Lambda_0|}{|\Lambda_n|}} \, \frac{\Gamma(a_n)}{\Gamma(a_0)} \, \frac{ {b_0}^{a_0}}{ {b_n}^{a_n}} = p(y|m) \; .\]

Thus, the log model evidence of this model is given by

\[\label{eq:GLM-NG-LME-s6} \begin{split} \log p(y|m) = \frac{1}{2} & \log |P| - \frac{n}{2} \log (2 \pi) + \frac{1}{2} \log |\Lambda_0| - \frac{1}{2} \log |\Lambda_n| + \\ & \log \Gamma(a_n) - \log \Gamma(a_0) + a_0 \log b_0 - a_n \log b_n \; . \end{split}\]
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Metadata: ID: P11 | shortcut: blr-lme | author: JoramSoch | date: 2020-01-03, 22:05.