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This question is about fitting a multivariate linear regression by maximum likelihood, under a specific parameterization of the covariance matrix, when the number of observations is smaller than the number of responses. It arises in an applied project that I'm part of.

Let $Y_i \in \mathbb R^r, i=1, \dots, n$ be independent multivariate normal with (non-stochastic) mean $\beta'X_i$ and covariance matrix $\Sigma = \Lambda (R_1 \otimes R_2)\Lambda$. $\Lambda$ is a diagonal matrix and the $R_i$ are correlation matrices of sizes $r_1$ and $r_2$, respectively, where $r_1 \times r_2 = r$.

The number of predictors, $p$, is small enough that the MLE of $\beta$ may be computed as $\hat{\beta} = (X'X)^{-1}X'Y$, where $Y$ and $X$ are $n\times r$ and $n\times p$ matrices with the $Y_i$ and $X_i$ as rows.

Now, after profiling out $\beta$ the profile log-likelihood is:

$$ \ell(\Lambda, R_1, R_2) = -\frac{nr}{2}\log(2\pi) - n\log\vert \Lambda\vert - \frac{nr_2}{2}\log\vert R_1\vert - \frac{nr_1}{2}\log\vert R_2\vert - \frac{n}{2}\mathrm{tr}\left[\Lambda^{-1}(R_1^{-1}\otimes R_2^{-1})\Lambda^{-1}S\right], $$

where $S = (Y - X\hat{\beta})'(Y - X\hat{\beta})/n$.

Question: Can this be optimized over $\Lambda, R_1, R_2$ subject to the constraint that $\Lambda$ is diagonal and $R_1,R_2$ are correlation matrices? I do have (unconstrained) gradients for all parameters.


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    $\begingroup$ Don't know how you computed gradient, and whether it is correct; presumably it does not account for constraints. So optimum may not have gradient = 0. You should solve as a constrained optimization problem (e.g., constrain diagonal elements of correlation matrices to 1, or don't even make them variables). Instead of Cholesky factorization type parameterization, consider imposing semidefintie constraints directly on the correlation matrices (constrain diagonals to 1) and solve with nonlinear SDP solver such as PENLAB. But don't really understand your problem specification or solution method. $\endgroup$ – Mark L. Stone Aug 4 '16 at 15:04
  • $\begingroup$ Thanks for your comments Mark. I skipped some details to make it shorter. I only expected the gradient of $\Lambda$ to be zero at solution since I left it unconstrained. I don't know how to impose PSD without Cholesky parameterization, but will have a look an PENALB. Thanks. $\endgroup$ – ekvall Aug 4 '16 at 15:09
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I have solved this problem using a blockwise coordinate descent-type algorithm. Given the low interest in this question I leave the details out but if anyone is interested in this or a similar problem and run into this question and answer I'd be happy to expand.

Repeat until convergence:

  1. Initialize $k = 0$, $\Lambda^0 = I$, $R_1^0 = I$ and $R_2^0 = I$
  2. Update $R_1^{k + 1}$ by solving the first order condition $\partial \ell(\Lambda^{k}, R_1, R_2^k)/\partial R_1^{-1} = 0$
  3. Update $R_2^{k + 1}$ by solving the first order condition $\partial \ell(\Lambda^{k}, R_1^{k + 1}, R_2)/\partial R_2^{-1} = 0$
  4. Rescale $R_1^{k + 1}$, $R_2^{k + 1}$ and $C^k$ to satisfy the constraints without changing the likelihood value. E.g., $C^k \leftarrow C^k[(R_1^{k + 1} \circ I) \otimes (R_2^{k + 1} \circ I)]^{1/2}$, where $\circ$ denotes Hadamard product and $\otimes$ denotes Kronecker product.
  5. Let $\Lambda = \mathrm{diag}(1/\theta_1, \dots, 1/\theta_m)$ and for $i = 1, \dots, m$ update $\theta_i$ by solving the first order condition $\partial \ell(\theta_1^{k + 1}, \dots, \theta_{i-1}^{k + 1}, \theta_i, \theta_{i + 1}^{k}, \dots \theta_m^k, R_1^{k +1}, R_2^{k + 1}) / \partial\theta_i = 0$
  6. $k \leftarrow k + 1$.

Some notes: every update is available in closed form, and every update is convex but the full problem is not convex in general, i.e. there is no guarantee that the point of convergence is a unique global maximum of the likelihood function.

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