Index: The Book of Statistical ProofsStatistical Models ▷ Univariate normal data ▷ Bayesian linear regression ▷ Combined posterior distribution from independent data sets

Theorem: Let $y = \left\lbrace y_1, \ldots, y_S \right\rbrace$ be a set of $S$ conditionally independent data sets assumed to follow linear regression models with design matrices $X_1, \ldots, X_S$, number of data points $n_1, \ldots, n_S$ and precision matrices $P_1, \ldots, P_n$, governed by identical regression coefficients $\beta$ and identical noise precision $\tau$:

$\label{eq:GLM-NG-S} \begin{split} y_1 &= X_1 \beta + \varepsilon_1, \; \varepsilon_1 \sim \mathcal{N}(0, \sigma^2 V_1), \; \sigma^2 V_1 = (\tau P_1)^{-1} \\ &\;\;\vdots \\ y_S &= X_S \beta + \varepsilon_S, \; \varepsilon_S \sim \mathcal{N}(0, \sigma^2 V_S), \; \sigma^2 V_S = (\tau P_S)^{-1} \; . \end{split}$

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 combined posterior distribution from observing these conditionally independent data sets is also given by a normal-gamma distribution

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

with the posterior hyperparameters

$\label{eq:GLM-NG-S-post-par} \begin{split} \mu_n &= \Lambda_n^{-1} \left( \sum_{i=1}^S X_i^\mathrm{T} P_i y_i + \Lambda_0 \mu_0 \right) \\ \Lambda_n &= \sum_{i=1}^S X_i^\mathrm{T} P_i X_i + \Lambda_0 \\ a_n &= a_0 + \frac{1}{2} \sum_{i=1}^S n_i \\ b_n &= b_0 + \frac{1}{2} \left( \sum_{i=1}^S y_i^\mathrm{T} P_i y_i + \mu_0^\mathrm{T} \Lambda_0 \mu_0 - \mu_n^\mathrm{T} \Lambda_n \mu_n \right) \; . \end{split}$

Proof: This can be seen by sequentially applying Bayes’ theorem for calculating the posterior distribution, while using the posterior after one iteration as the prior for the next iteration.

Let $\mu_0^{(i)}, \Lambda_0^{(i)}, a_0^{(i)}, b_0^{(i)}$ denote the prior hyperparameters before analyzing the $i$-th data set, such that e.g. $\mu_0^{(1)}$ is identical to $\mu_0$ in \eqref{eq:GLM-NG-prior}:

$\label{eq:GLM-NG-S-prior-y1} \begin{split} \mu_0^{(1)} &= \mu_0 \\ \Lambda_0^{(1)} &= \Lambda_0 \\ a_0^{(1)} &= a_0 \\ b_0^{(1)} &= b_0 \; . \end{split}$

Moreover, let $\mu_n^{(i)}, \Lambda_n^{(i)}, a_n^{(i)}, b_n^{(i)}$ denote the posterior hyperparameters after analyzing the $i$-th data set, such that e.g. $\mu_n^{(S)}$ is identical to $\mu_n$ in \eqref{eq:GLM-NG-post}:

$\label{eq:GLM-NG-S-post-yS} \begin{split} \mu_n^{(S)} &= \mu_n \\ \Lambda_n^{(S)} &= \Lambda_n \\ a_n^{(S)} &= a_n \\ b_n^{(S)} &= b_n \; . \end{split}$

The posterior after seeing the $i$-th data set is equal to the prior before seeing the $(i+1)$-th data set, so we have the relation:

$\label{eq:GLM-NG-S-prior-post} \begin{split} \mu_0^{(i+1)} &= \mu_n^{(i)} \\ \Lambda_0^{(i+1)} &= \Lambda_n^{(i)} \\ a_0^{(i+1)} &= a_n^{(i)} \\ b_0^{(i+1)} &= b_n^{(i)} \; . \end{split}$

The posterior distribution for Bayesian linear regression when observing a single data set is given by the following hyperparameter equations:

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

We can apply \eqref{eq:GLM-NG-post-par} to calculate the posterior hyperparameters after seeing the first data set:

$\label{eq:GLM-NG-S-post-y1} \begin{split} \mu_n^{(1)} &= {\Lambda_n^{(1)}}^{-1} \left( X_1^\mathrm{T} P_1 y_1 + \Lambda_0^{(1)} \mu_0^{(1)} \right) \\ &= {\Lambda_n^{(1)}}^{-1} \left( X_1^\mathrm{T} P_1 y_1 + \Lambda_0 \mu_0 \right) \\ \Lambda_n^{(1)} &= X_1^\mathrm{T} P_1 X_1 + \Lambda_0^{(1)} \\ &= X_1^\mathrm{T} P_1 X_1 + \Lambda_0 \\ a_n^{(1)} &= a_0^{(1)} + \frac{1}{2} n_1 \\ &= a_0 + \frac{1}{2} n_1 \\ b_n^{(1)} &= b_0^{(1)} + \frac{1}{2} \left( y_1^\mathrm{T} P_1 y_1 + {\mu_0^{(1)}}^\mathrm{T} \Lambda_0^{(1)} \mu_0^{(1)} - {\mu_n^{(1)}}^\mathrm{T} \Lambda_n^{(1)} \mu_n^{(1)} \right) \\ &= b_0 + \frac{1}{2} \left( y_1^\mathrm{T} P_1 y_1 + \mu_0^\mathrm{T} \Lambda_0 \mu_0 - {\mu_n^{(1)}}^\mathrm{T} \Lambda_n^{(1)} \mu_n^{(1)} \right) \; . \end{split}$

These are the prior hyperparameters before seeing the second data set:

$\label{eq:GLM-NG-S-prior-y2} \begin{split} \mu_0^{(2)} &= \mu_n^{(1)} \\ \Lambda_0^{(2)} &= \Lambda_n^{(1)} \\ a_0^{(2)} &= a_n^{(1)} \\ b_0^{(2)} &= b_n^{(1)} \; . \end{split}$

Thus, we can again use \eqref{eq:GLM-NG-post-par} to calculate the posterior hyperparameters after seeing the second data set:

$\label{eq:GLM-NG-S-post-y2} \begin{split} \mu_n^{(2)} &= {\Lambda_n^{(2)}}^{-1} \left( X_2^\mathrm{T} P_2 y_2 + \Lambda_0^{(2)} \mu_0^{(2)} \right) \\ &= {\Lambda_n^{(2)}}^{-1} \left( X_2^\mathrm{T} P_2 y_2 + \Lambda_n^{(1)} {\Lambda_n^{(1)}}^{-1} \left( X_1^\mathrm{T} P_1 y_1 + \Lambda_0 \mu_0 \right) \right) \\ &= {\Lambda_n^{(2)}}^{-1} \left( X_1^\mathrm{T} P_1 y_1 + X_2^\mathrm{T} P_2 y_2 + \Lambda_0 \mu_0 \right) \\ \Lambda_n^{(2)} &= X_2^\mathrm{T} P_2 X_2 + \Lambda_0^{(2)} \\ &= X_2^\mathrm{T} P_2 X_2 + X_1^\mathrm{T} P_1 X_1 + \Lambda_0 \\ &= X_1^\mathrm{T} P_1 X_1 + X_2^\mathrm{T} P_2 X_2 + \Lambda_0 \\ a_n^{(2)} &= a_0^{(2)} + \frac{1}{2} n_2 \\ &= a_0 + \frac{1}{2} n_1 + \frac{1}{2} n_2 \\ &= a_0 + \frac{1}{2} \left( n_1 + n_2 \right) \\ b_n^{(2)} &= b_0^{(2)} + \frac{1}{2} \left( y_2^\mathrm{T} P_2 y_2 + {\mu_0^{(2)}}^\mathrm{T} \Lambda_0^{(2)} \mu_0^{(2)} - {\mu_n^{(2)}}^\mathrm{T} \Lambda_n^{(2)} \mu_n^{(2)} \right) \\ &= b_0 + \frac{1}{2} \left( y_1^\mathrm{T} P_1 y_1 + \mu_0^\mathrm{T} \Lambda_0 \mu_0 - {\mu_n^{(1)}}^\mathrm{T} \Lambda_n^{(1)} \mu_n^{(1)} \right) + \frac{1}{2} \left( y_2^\mathrm{T} P_2 y_2 + {\mu_n^{(1)}}^\mathrm{T} \Lambda_n^{(1)} \mu_n^{(1)} - {\mu_n^{(2)}}^\mathrm{T} \Lambda_n^{(2)} \mu_n^{(2)} \right) \\ &= b_0 + \frac{1}{2} \left( y_1^\mathrm{T} P_1 y_1 + y_2^\mathrm{T} P_2 y_2 + \mu_0^\mathrm{T} \Lambda_0 \mu_0 - {\mu_n^{(2)}}^\mathrm{T} \Lambda_n^{(2)} \mu_n^{(2)} \right) \; . \end{split}$

These are the prior hyperparameters before seeing the third data set:

$\label{eq:GLM-NG-S-prior-y3} \begin{split} \mu_0^{(3)} &= \mu_n^{(2)} \\ \Lambda_0^{(3)} &= \Lambda_n^{(2)} \\ a_0^{(3)} &= a_n^{(2)} \\ b_0^{(3)} &= b_n^{(2)} \; . \end{split}$

Generalizing this, we have after observing the $j$-th data set:

$\label{eq:GLM-NG-S-post-yi} \begin{split} \mu_n^{(j)} &= {\Lambda_n^{(j)}}^{-1} \left( \sum_{i=1}^j X_i^\mathrm{T} P_i y_i + \Lambda_0 \mu_0 \right) \\ \Lambda_n^{(j)} &= \sum_{i=1}^j X_i^\mathrm{T} P_i X_i + \Lambda_0 \\ a_n^{(j)} &= a_0 + \frac{1}{2} \sum_{i=1}^j n_i \\ b_n^{(j)} &= b_0 + \frac{1}{2} \left( \sum_{i=1}^j y_i^\mathrm{T} P_i y_i + \mu_0^\mathrm{T} \Lambda_0 \mu_0 - {\mu_n^{(j)}}^\mathrm{T} \Lambda_n^{(j)} \mu_n^{(j)} \right) \; . \end{split}$

Plugging in $j = S$, we obtain the final posterior distribution:

$\label{eq:GLM-NG-S-post-par-qed} \begin{split} \mu_n = \mu_n^{(S)} &= {\Lambda_n^{(S)}}^{-1} \left( \sum_{i=1}^S X_i^\mathrm{T} P_i y_i + \Lambda_0 \mu_0 \right) = \Lambda_n^{-1} \left( \sum_{i=1}^S X_i^\mathrm{T} P_i y_i + \Lambda_0 \mu_0 \right) \\ \Lambda_n = \Lambda_n^{(S)} &= \sum_{i=1}^S X_i^\mathrm{T} P_i X_i + \Lambda_0 \\ a_n = a_n^{(S)} &= a_0 + \frac{1}{2} \sum_{i=1}^S n_i \\ b_n = b_n^{(S)} &= b_0 + \frac{1}{2} \left( \sum_{i=1}^S y_i^\mathrm{T} P_i y_i + \mu_0^\mathrm{T} \Lambda_0 \mu_0 - {\mu_n^{(S)}}^\mathrm{T} \Lambda_n^{(S)} \mu_n^{(S)} \right) \\ &= b_0 + \frac{1}{2} \left( \sum_{i=1}^S y_i^\mathrm{T} P_i y_i + \mu_0^\mathrm{T} \Lambda_0 \mu_0 - \mu_n^\mathrm{T} \Lambda_n \mu_n \right) \; . \end{split}$

This result is also compatible with the general theorem about combined posterior distributions in terms of individual posterior distributions when analyzing independent data sets.

Sources:

Metadata: ID: P447 | shortcut: blr-postind | author: JoramSoch | date: 2024-04-12, 16:31.