Efficient factoring of GP likelihoods


There are many ways to cleverly slice up GP likelihoods so that inference is cheap.

This page is about some of them, especially the union of sparse and variational tricks. Scalable Gaussian process regressions choose cunning factorisations such that the model collapses down to a lower-dimensional thing than it might have seemed to need, at least approximately. There is a comptilation of tricks to make this go — variational approximations a model, sparse GP models where there are a small number of inducing points (Dezfouli and Bonilla 2015; Edwin V. Bonilla, Krauth, and Dezfouli 2019; Krauth et al. 2016; Hensman, Fusi, and Lawrence 2013; Salimbeni and Deisenroth 2017). You might suspect yourself of using such a method if you find that some important high-dimensional expectation can be evaluated by some function of univariate Gaussians.

This is a related notion to other tricks which factorise a distribution cleverly, such message-passing inference. There are indeed a lot of different factorisations that can be done here; See filtering GPs for one which factorizes over a single input axis. Also Toeplitz and related structures work out nicely for, e.g. lattice-distributed inputs and some other situations I forget right now. I will bet you they can all be used together.

Basic sparsity via inducing variables

Sparse Gaussian processes are ones where you approximate the target posterior by summarising the data with a short list of inducing points that have nearly the same similar posterior density, by some metric. These have been invented in various forms by various people, but most of the differences we can ignore. Based on citations, thethe Right Way is the version of Titsias (2009), which we use in all the subsequent things as a building-block.

Here is my explanation, based on Hensman, Fusi, and Lawrence (2013) summarizing Titsias (2009).

We consider an output vector \(\mathbf{y},\) where each entry \(y_{i}\) is a noisy observation of the function \(f,\) i.e. \(y_{i}=f(\mathbf{x}_{i})+\varepsilon_{i}\) for all the points \(\mathbf{X}=\left\{\mathbf{x}_{i}\right\}_{i=1}^{n}.\) We assume the noises \(\varepsilon_{i}\sim\mathcal{N}(0,1/\beta)\) to be independent. We introduce a Gaussian process prior over \(f(\cdot).\) Now let the vector \(\mathbf{f}=\{f(x_{i})\}_{i=1}^{n}\) contain values of the function evaluated at the inputs. We introduce a set of inducing variables, defined similarly: let the vector \(\mathbf{u}=\{f(\mathbf{z}_{i})\}_{i=1}^{m}\). Standard properties of Gaussian processes allow us to write \[ \begin{aligned} p(\mathbf{y} \mid \mathbf{f}) &=\mathcal{N}\left(\mathbf{y} \mid \mathbf{f}, \beta^{-1} \mathbf{I}\right) \\ p(\mathbf{f} \mid \mathbf{u}) &=\mathcal{N}\left(\mathbf{f} \mid \mathbf{K}_{n m} \mathbf{K}_{m m}^{-1} \mathbf{u}, \tilde{\mathbf{K}}\right) \\ p(\mathbf{u}) &=\mathcal{N}\left(\mathbf{u} \mid \mathbf{0}, \mathbf{K}_{m m}\right) \end{aligned} \] where \(\mathbf{K}_{m m}\) is the covariance function evaluated between all the inducing points and \(\mathbf{K}_{n m}\) is the covariance function between all inducing points and training points and we have defined with \(\tilde{\mathbf{K}}=\mathbf{K}_{n n}-\) \(\mathbf{K}_{n m} \mathbf{K}_{m m}^{-1} \mathbf{K}_{m n}\)

Quoting them:

We first apply Jensen’s inequality on the conditional probability \(p(\mathbf{y} \mid \mathbf{u})\) \[ \begin{aligned} \log p(\mathbf{y} \mid \mathbf{u}) &=\log \langle p(\mathbf{y} \mid \mathbf{f})\rangle_{p(\mathbf{f} \mid \mathbf{u})} \\ & \geq\langle\log p(\mathbf{y} \mid \mathbf{f})\rangle_{p(\mathbf{f} \mid \mathbf{u})} \triangleq \mathcal{L}_{1} \end{aligned} \] where \(\langle\cdot\rangle_{p(x)}\) denotes an expectation under \(p(x).\) For Gaussian noise taking the expectation inside the log is tractable, but it results in an expression containing \(\mathbf{K}_{n n}^{-1},\) which has a computational complexity of \(\mathcal{O}\left(n^{3}\right).\) Bringing the expectation outside the log gives a lower bound, \(\mathcal{L}_{1},\) which can be computed with has complexity \(\mathcal{O}\left(m^{3}\right).\) Further, when \(p(\mathbf{y} \mid \mathbf{f})\) factorises across the data, \[ p(\mathbf{y} \mid \mathbf{f})=\prod_{i=1}^{n} p\left(y_{i} \mid f_{i}\right) \] then this lower bound can be shown to be separable across y giving \[ \exp \left(\mathcal{L}_{1}\right)=\prod_{i=1}^{n} \mathcal{N}\left(y_{i} \mid \mu_{i}, \beta^{-1}\right) \exp \left(-\frac{1}{2} \beta \tilde{k}_{i, i}\right) \] where \(\boldsymbol{\mu}=\mathbf{K}_{n m} \mathbf{K}_{m m}^{-1} \mathbf{u}\) and \(\tilde{k}_{i, i}\) is the \(i\) th diagonal element of \(\widetilde{\mathbf{K}}\). Note that the difference between our bound and the original log likelihood is given by the Kullback Leibler (KL) divergence between the posterior over the mapping function given the data and the inducing variables and the posterior of the mapping function given the inducing variables only, \[ \mathrm{KL}(p(\mathbf{f} \mid \mathbf{u}) \| p(\mathbf{f} \mid \mathbf{u}, \mathbf{y})) \] This KL divergence is minimized when there are \(m=n\) inducing variables and they are placed at the training data locations. This means that \(\mathbf{u}=\mathbf{f}\), \(\mathbf{K}_{m m}=\mathbf{K}_{n m}=\mathbf{K}_{n n}\) meaning that \(\tilde{\mathbf{K}}=\mathbf{0} .\) In this case we recover \(\exp \left(\mathcal{L}_{1}\right)=p(\mathbf{y} \mid \mathbf{f})\) and the bound becomes equality because \(p(\mathbf{f} \mid \mathbf{u})\) is degenerate. However, since \(m=n\) and that there would be no computational or storage advantage from the representation. When \(m<n\) the bound can be maximised with respect to \(\mathbf{Z}\) (which are variational parameters). This minimises the KL divergence and ensures that \(\mathbf{Z}\) are distributed amongst the training data \(\mathbf{X}\) such that all \(\tilde{k}_{i, i}\) are small. In practice this means that the expectations in (1) are only taken across a narrow domain (\(\tilde{k}_{i, i}\) is the marginal variance of \(p\left(f_{i} \mid \mathbf{u}\right)\) ), keeping Jensen’s bound tight.

…we recover the bound of Titsias (2009) by marginalising the inducing variables, \[ \begin{aligned} \log p(\mathbf{y} \mid \mathbf{X}) &=\log \int p(\mathbf{y} \mid \mathbf{u}) p(\mathbf{u}) \mathrm{d} \mathbf{u} \\ & \geq \log \int \exp \left\{\mathcal{L}_{1}\right\} p(\mathbf{u}) \mathrm{d} \mathbf{u} \triangleq \mathcal{L}_{2} \end{aligned} \] which with some linear algebraic manipulation leads to \[ \mathcal{L}_{2}=\log \mathcal{N}\left(\mathbf{y} \mid \mathbf{0}, \mathbf{K}_{n m} \mathbf{K}_{m m}^{-1} \mathbf{K}_{m n}+\beta^{-1} \mathbf{I}\right)-\frac{1}{2} \beta \operatorname{tr}(\widetilde{\mathbf{K}}) \] matching the result of Titsias, with the implicit approximating distribution \(q(\mathbf{u})\) having precision \[ \mathbf{\Lambda}=\beta \mathbf{K}_{m m}^{-1} \mathbf{K}_{m n} \mathbf{K}_{n m} \mathbf{K}_{m m}^{-1}+\mathbf{K}_{m m}^{-1} \] and mean \[ \hat{\mathbf{u}}=\beta \mathbf{\Lambda}^{-1} \mathbf{K}_{m m}^{-1} \mathbf{K}_{m n} \mathbf{y} \]

SVI for Gaussian processes

As seen in Hensman, Fusi, and Lawrence (2013);Salimbeni and Deisenroth (2017).

Latent Gaussian Process models

Here is the Edwin V. Bonilla, Krauth, and Dezfouli (2019) set up for Latent Gaussian Process models (“LGPMs”).

We are learning a mapping \(\boldsymbol{f}:\mathbb{R}^D\to\mathbb{R}^P\) from data. The dataset looks like \(\mathcal{D}=\left\{\mathbf{x}_{n}, \mathbf{y}_{n}\right\}_{n=1}^{N}\equiv \left\{\mathbf{x}, \mathbf{y}\right\}.\) \(\mathbf{x}_{n}\in \mathbb{R}^D\) is an input vector and \(\mathbf{y}_{n}\in\mathbb{R}^P\) is an output. We decree that the mapping from inputs to outputs, may be expressed by \(Q\) underlying latent functions \(\left\{f_{j}\right\}_{j=1}^{Q}.\) We assume that the \(Q\) latent functions \(\left\{f_{j}\right\}\) are drawn from (a priori) independent zero-mean Gaussian processes.

\[ \begin{aligned} p\left(f_{j} \mid \boldsymbol{\theta}_{j}\right) & \sim \mathcal{G} \mathcal{P}\left(0, \kappa_{j}\left(\cdot, \cdot ; \boldsymbol{\theta}_{j}\right)\right), \quad j=1, \ldots Q, \quad \text { and } \\ p(\mathbf{f} \mid \boldsymbol{\theta}) &=\prod_{j=1}^{Q} p\left(\mathbf{f}_{\cdot j} \mid \boldsymbol{\theta}_{j}\right) \\ &=\prod_{j=1}^{Q} \mathcal{N}\left(\mathbf{f}_{\cdot j} ; \mathbf{0}, \mathbf{K}_{\mathbf{x x}}^{j}\right). \end{aligned} \] Here \(\mathbf{f}\) is the set of all latent function values; \(\mathbf{f}_{\cdot j}=\left\{f_{j}\left(\mathbf{x}_{n}\right)\right\}_{n=1}^{N}\) denotes the values of latent function \(j\). The Gram matrix is \(\mathbf{K}_{\mathrm{xx}}^{j}\), induced by a covariance kernel, \(\kappa_{j}\left(\cdot, \cdot ; \boldsymbol{\theta}_{j}\right)\). The parameters of all kernel functions we call \(\boldsymbol{\theta}=\left\{\boldsymbol{\theta}_{j}\right\}.\) Our observation model can have various likelihoods; We will call the corresponding parameter \(\boldsymbol{\phi}\). We assume that our multi-dimensional observations \(\left\{\mathbf{y}_{n}\right\}\) are i.i.d. given the latent functions \(\left\{\mathbf{f}_{n}\right\},\) so that \[ p(\mathbf{y} \mid \mathbf{f}, \boldsymbol{\phi})=\prod_{n=1}^{N} p\left(\mathbf{y}_{n} \mid \mathbf{f}_{n \cdot}, \boldsymbol{\phi}\right) \] \(\mathbf{f}_{n\cdot}=\{f_{j}(\boldsymbol{x}_n)\}_{j=1}^{q}\) is the set of latent \(\boldsymbol{f}\) values upon which \(\mathbf{y}_{n}\) depends.

There are several factorizations to note here

  1. The prior is factored into latent functions per-coordinate
  2. the conditional likelihood is factored over observations (i.e. nosie in independent)
  3. we factorise the variational approximating distribution into mixtures (TBD)

This is going to work out well for us when we try to devise a system of inference later to minimise the ELBO. TBC.

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