Index: The Book of Statistical ProofsStatistical Models ▷ Univariate normal data ▷ Bayesian linear regression ▷ Maximum-a-posteriori estimation

Theorem: Let there be a linear regression model

$\label{eq:GLM} y = X\beta + \varepsilon, \; \varepsilon \sim \mathcal{N}(0, \sigma^2 V), \; \sigma^2 V = (\tau P)^{-1}$

and 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 maximum-a-posteriori estimates of $\beta$ and $\tau$ are

$\label{eq:GLN-NG-MAP} \begin{split} \hat{\beta}_\mathrm{MAP} &= (X^\mathrm{T} P X + \Lambda_0)^{-1} (X^\mathrm{T} P y + \Lambda_0 \mu_0) \\ \hat{\tau}_\mathrm{MAP} &= \left( 2 a_0 + n - 2 \right) \left( 2 b_0 + (y - X \hat{\beta}_\mathrm{MAP})^\mathrm{T} P (y - X \hat{\beta}_\mathrm{MAP}) + (\hat{\beta}_\mathrm{MAP} - \mu_0)^\mathrm{T} \Lambda_0 (\hat{\beta}_\mathrm{MAP} - \mu_0) \right)^{-1} \end{split}$

where $n$ is the number of data points.

Proof: Given the prior distribution in \eqref{eq:GLM-NG-prior}, the posterior distribution for multiple linear regression is also a normal-gamma distribution

$\label{eq:GLM-NG-post} p(\beta,\tau|y) = \mathcal{N}(\beta; \mu_n, (\tau \Lambda_n)^{-1}) \cdot \mathrm{Gam}(\tau; a_n, b_n)$

where the posterior hyperparameters are equal to

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

From this, the conditional posterior distribution over $\beta$ follows as

$\label{eq:GLM-NG-post-beta} p(\beta|\tau,y) = \mathcal{N}(\beta; \mu_n, (\tau \Lambda_n)^{-1})$

and the marginal posterior distribution over $\tau$ follows as

$\label{eq:GLM-NG-post-tau} p(\tau|y) = \mathrm{Gam}(\tau; a_n, b_n) \; .$

The mode of the multivariate normal distribution is given by

$\label{eq:mvn-mode} X \sim \mathcal{N}(\mu, \Sigma) \quad \Rightarrow \quad \mathrm{mode}(X) = \mu$

and the mode of the gamma distribution is given by

$\label{eq:gam-mode} X \sim \mathrm{Gam}(a, b) \quad \Rightarrow \quad \mathrm{mode}(X) = \frac{a-1}{b} \; .$

Applying \eqref{eq:mvn-mode} to \eqref{eq:GLM-NG-post-beta}, the maximum-a-posteriori estimate of $\beta$ follows as

$\label{eq:GLN-NG-MAP-beta} \begin{split} \hat{\beta}_\mathrm{MAP} &= \mu_n \\ &= \Lambda_n^{-1} (X^\mathrm{T} P y + \Lambda_0 \mu_0) \\ &= (X^\mathrm{T} P X + \Lambda_0)^{-1} (X^\mathrm{T} P y + \Lambda_0 \mu_0) \end{split}$

and applying \eqref{eq:gam-mode} to \eqref{eq:GLM-NG-post-tau}, the maximum-a-posteriori estimate of $\tau$ follows as

$\label{eq:GLN-NG-MAP-tau} \begin{split} \hat{\tau}_\mathrm{MAP} &= \frac{a_n-1}{b_n} \\ &= \left( a_0 + \frac{n}{2} - 1 \right) \left( b_0 + \frac{1}{2} (y^\mathrm{T} P y + \mu_0^\mathrm{T} \Lambda_0 \mu_0 - \mu_n^\mathrm{T} \Lambda_n \mu_n) \right)^{-1} \\ &= \left( 2 a_0 + n - 2 \right) \left( 2 b_0 + y^\mathrm{T} P y + \mu_0^\mathrm{T} \Lambda_0 \mu_0 - \mu_n^\mathrm{T} \Lambda_n \mu_n \right)^{-1} \\ &= \left( 2 a_0 + n - 2 \right) \left( 2 b_0 + y^\mathrm{T} P y + \mu_0^\mathrm{T} \Lambda_0 \mu_0 - \hat{\beta}_\mathrm{MAP}^\mathrm{T} \left( X^\mathrm{T} P X + \Lambda_0 \right) \hat{\beta}_\mathrm{MAP} \right)^{-1} \\ &= \left( 2 a_0 + n - 2 \right) \left( 2 b_0 + y^\mathrm{T} P y + \mu_0^\mathrm{T} \Lambda_0 \mu_0 - \hat{\beta}_\mathrm{MAP}^\mathrm{T} X^\mathrm{T} P X \hat{\beta}_\mathrm{MAP} - \hat{\beta}_\mathrm{MAP}^\mathrm{T} \Lambda_0 \hat{\beta}_\mathrm{MAP} \right)^{-1} \\ &= \left( 2 a_0 + n - 2 \right) \left( 2 b_0 + (y - X \hat{\beta}_\mathrm{MAP})^\mathrm{T} P (y - X \hat{\beta}_\mathrm{MAP}) + (\hat{\beta}_\mathrm{MAP} - \mu_0)^\mathrm{T} \Lambda_0 (\hat{\beta}_\mathrm{MAP} - \mu_0) \right)^{-1} \; . \end{split}$
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Metadata: ID: P448 | shortcut: blr-map | author: JoramSoch | date: 2024-04-19, 14:45.