The EM GLM algorithm - Proof of predicted values

Robert M. Cook

2019-07-01

The EM algorithm treats the observed data as arising from one of several competing random processes. Each process has some underpinning probability distribution:

\[P(y_i | x_i, B_i)\]

and likelihood of all observations:

\[ L(Y | X, B_1 ... B_k) = \prod_{i=1}^N \prod_{j=1}^k P(y_i | x_i, B_j )^{I(Z_i = j)} \]

where each observation has a hidden variable, \(Z_i\) (indicating which of the competing random processes gives rise to the observation) and the function \(I(Z_i = j)\) is an indicator function (i.e. it equals 1 when true and 0 otherwise). If \(Z_i\) were known the observations could be grouped and \(k\) models fit so as to maximize the log-likelihood:

\[ l(Y | X, B_1 ... B_k) = \sum_{i=1}^N \sum_{j=1}^k I(Z_i = j) \ln(P(y_i | x_i, B_j )) \]

Given that \(Z_i\) is not known, we instead substitute \(Z_i\) for a probability \(T_{j}\). \(T_{j}\) is the probability of a given observation arising from one of the \(k\) competing models (assuming the proposed parameters are true) normalized across all models. Making this substitution, the likelihood of the \(i^{th}\) observation, \(l_i\) is:

\[ l_i(y | x, B_1 ... B_k, T_1 ... T_k) = \sum_{j=1}^k T_{j} \ln(P(y | x, B_j)) \]

with \(\sum_{j=1}^k T_{i,j} = 1\). For proposed parameters, \(B_1 ... B_k\), \(T_j\) has values:

\[ T_{j} = \frac{P(y | x, B_j)}{\sum_{j=1}^k P(y | x, B_k) } \]

Using this replacement of \(I(Z_i = j)\) with \(T_{j}\) each observation becomes a mixed distribution:

\[P(y| x, B_1 .. B_K, T_1 ... T_k) = \alpha \prod_{j=1}^k P(y | x, B_j) ^ {T_{j}}\]

with \(\alpha\) being some additional normalization term.

Binomial example:

For the Binomial distribution the mixed probability mass function is:

\[ P(y | p_1 ... p_k, T_1 ..T_k) = f_{norm.}(n, p1 ... p_k) \prod_{i=1}^k \left({n \choose y} p_i^y(1-p)^{n-y}\right)^{T_i} \]

where \(f_{norm.}\) is a normalization coefficient to be found.

Re-arranging the powers we can re-write the equation as:

\[ P(y | p_1 ... p_k, T_1 ..T_k) = f_{norm.}(n, p_1.. p_n, T_1 ... T_n) {n \choose y} \left( \prod_{i=1}^k p_i^{T_i} \right)^y \left(\prod_{i=1}^k (1-p_i)^{T_i}\right)^{n - y} \]

Which suddenly resembles the standard binomial process. Using the binomial theorem we know that:

\[\sum_{y=0}^n {n \choose y} \left( \prod_{i=1}^k p_i^{T_i} \right)^y \left(\prod_{i=1}^k (1-p_i)^{T_i}\right)^{n - y} = (\prod_{i=1}^k p_i^{T_i} + \prod_{i=1}^k (1-p_i)^{T_i})^n \]

and hence we have the normalization term:

\[f_{norm.}(n, p_1.. p_n, T_1 ... T_n) = (\prod_{i=1}^k p_i^{T_i} + \prod_{i=1}^k (1-p_i)^{T_i})^{-n}\]

And our original expression for \(P(y)\) can be rewritten as:

\[ P(y | p_1 ... p_k, T_1 ..T_k) = {n \choose y} \frac{\left( \prod_{i=1}^k p_i^{T_i} \right)^y \left(\prod_{i=1}^k (1-p_i)^{T_i}\right)^{n - y}}{(\prod_{i=1}^k p_i^{T_i} + \prod_{i=1}^k (1-p_i)^{T_i})^{n}}\]

From here we can re-write in terms of \(p_{pooled}\) where:

\[p_{pooled} = \frac{\prod_{i=1}^k p_i^{T_i} }{\prod_{i=1}^k p_i^{T_i} + \prod_{i=1}^k (1-p_i)^{T_i}} \]

\[(1 - p_{pooled}) = \frac{\prod_{i=1}^k (1 - p_i)^{T_i} }{\prod_{i=1}^k p_i^{T_i} + \prod_{i=1}^k (1-p_i)^{T_i}}\]

\[ P(y| p_{pooled}) = {n \choose y} p_{pooled}^y (1- p_{pooled})^{n - y} \]

now - if we use the canonical link function (such that \(\theta = xB\)) we have:

\[\theta_{pooled} = log\left(\frac{p_{pooled}}{1 - p_{pooled}}\right) = log\left(\frac{\prod_{i=1}^k p_i^{T_i} }{\prod_{i=1}^k ( 1- p_i)^{T_i} }\right) = log\left( \prod_{i=1}^k \left( \frac{p_i}{1-p_i}\right)^{T_i} \right) = \sum_{i=1}^k T_i \theta_i = \sum_{i=1}^k T_i x B_i \]

From here the standard rules of GLMs can be applied, with \(\mu = b'(\theta_{pooled}).\)

Poisson derivation

For the Poisson distribution the mixed probability mass function is:

\[ P(y |\lambda_1 ... \lambda_k, T_1 ..T_k) = f_{norm.}(n, \lambda_1 ... \lambda_k) \prod_{i=1}^k \left(\frac{\lambda_i^y}{y!}\right)^{T_i} \]

where \(f\) is again the normalization function. Following the same approach as above, we can re-arrange the powers:

\[ P(y |\lambda_1 ... \lambda_k, T_1 ..T_k) = f_{norm.}(n,\lambda_1 ... \lambda_k) \frac{\left( \prod_{i=1}^k \lambda_i^{T_i} \right)^y }{y!} \]

which follows the same form as the typical Poisson dist. with a value of:

\[ \lambda_{pooled} = \prod_{i=1}^k \lambda_i^{T_i} \]

\[ P(y | \lambda_1 ... \lambda_k, T_1 ..T_k) = \frac{e^{-\lambda_{pooled}} \lambda_{pooled}}{y!} \]

Now, we can again apply the canonical link function, \(\theta_i = log(\lambda_i)\) and \(\lambda_i = e^{\theta_i}\) with \(\theta_i = xB_i\):

\[\theta_{pooled} = \log(\lambda_{pooled}) = \log \left(\prod_{i=1}^k \lambda_i^{T_i} \right) = \sum_{i=1}^k T_i \log(\lambda_i) =\sum_{i=1}^k T_i x B_i \]

Discussion

By using the canonical link both the Poisson and Binomial distribution result in the same link between \(\theta_{pooled}\) and the EM_GLM parameters \(B_i\).

With this proof, we can then assume that the standard GLM results from \(b(\theta)\) will hold, simplifying prediction and residuals.