The interpretation of densities as intensities and vice versa

Basis function method for density

Let’s consider the case where we try to estimate this function by constructing it from some given basis of \(p\) functions \(\phi_j: \mathbb{R}^d\rightarrow[0,\infty),\) so that

\[f(x)=\sum_{j\leq p}w_j\phi_j(x)\]

and \(\theta\equiv\{w_j\}_{j\leq p}.\) We keep this simple by requiring \(\int\phi_j(x)dx=1,\) so that they are all valid densities. Then the requirement that \(\int f(x)dx=1\) will imply that \(\sum_j w_j=1,\) i.e. we are taking a convex combination of these basis densities.

Then the maximum likelihood estimator can be written

\[ \begin{aligned} \hat{\theta} &=\operatorname{argmax}_{\{w_i\}}f(\{x_i\};\{w_i\})\\ &=\operatorname{argmax}_{\{w_i\}}\sum_i \log \sum_{j\leq p}w_j\phi_j(x_i) \end{aligned} \]

A moment’s thought reveals that this equation has no solution, since it is strictly increasing in each \(w_j\). However, we are missing a constraint, which is that to be a well-defined probability density, it must integrate to unity, i.e.

\[ \int f(\{x\};\{w_i\})dx = 1 \] and therefore

\[ \begin{aligned} \int \sum_{j\leq p}w_j\phi_j(x)dx&=1\\ \sum_{j\leq p}w_j\int\phi_j(x)dx&=1\\ \sum_{j\leq p}w_j &=1\\ \end{aligned} \]

By messing around with Lagrange multipliers to enforce that constraint we eventually find

\[ \hat{\{w_k\}} = \frac{\sum_i \phi_k(x_i)}{\sum_i \sum_j \phi_j(x_i)} \]

Intensities

Consider the problem of estimating the intensity \(\lambda\) of a simple, non-interacting inhomogeneous point process \(N(B)\) on some compact \(W\subset\mathbb{R}^d\) from a realisation \(\{x_i\}_{i\leq n}\), and this counting function \(N(B)\) counts the number of points that fall on a set \(B\subset\mathbb{R}^d\).

The intensity is (in the simple non interacting case — see Daley and Vere-Jones (2003) for other cases) a function \(\lambda:\mathbb{R}\rightarrow [0,\infty)\) such that, or any box \(B\subset W\),

\[N(B)\sim\operatorname{Poisson}(\Lambda(B))\] where

\[\Lambda(B):=\int_Bd\lambda(x)dx\] and for any disjoint boxes, \(N(A)\perp N(B).\)

After some argumentation about intensities we can find a likelihood for the observed distribution:

\[f(\{x_i\};\tau)= \prod_i \lambda(x_i;\tau)\exp\left(-\int\lambda(x;\tau)dx\right). \]

Say that we wish to find the inhomogeneous intensity function by the method of maximum likelihood. We allow the intensity function to be described by a parameter vector \(\tau,\) which we write \(\lambda(x;\tau)\), and we once again construct an estimate:

\[ \begin{aligned} \hat{\tau}&:=\operatorname{argmax}_\tau\sum_i\log f(x;\tau)\\ &=\operatorname{argmax}_\tau\sum_i\log \left(\lambda(x_i;\tau) \exp\left(-\int_W\lambda(x;\tau) dx\right)\right)\\ &=\operatorname{argmax}_\tau\sum_i\log \lambda(x_i;\tau)-\int_W\lambda(x;\tau)dx\\ &=\operatorname{argmax}_\tau\sum_i\log \lambda(x_i;\tau)-\Lambda(W) \end{aligned} \]

Basis function method for intensity

Now consider the case where we assume that the intensity can be written in a \(\phi_k\) basis as above, so that

\[\lambda(x)=\sum_{j\leq p}\omega_j\phi_j(x)\]

with \(\tau\equiv\{\omega_j\}.\) Then our estimate may be written

\[ \begin{aligned} \hat{\tau}&:=\operatorname{argmax}_{\{\omega_j\}}f\left(\{x_i\};\{\omega_j\} \right)\\ &:=\operatorname{argmax}_{\{\omega_j\}}\sum_i\left(\log \lambda(x_i;\tau)-\Lambda(W)\right)\\ &=\operatorname{argmax}_{\{\omega_j\}}\sum_i\left(\log \sum_{j\leq p}\omega_j\phi_j(x_i)-\int_W\sum_{j\leq p}\omega_j\phi_j(x)dx\right)\\ &=\operatorname{argmax}_{\{\omega_j\}}\sum_i\left(\log \sum_{j\leq p}\omega_j\phi_j(x_i)-\sum_{j\leq p}\omega_j\int_W \phi_j(x)dx \right)\\ &=\operatorname{argmax}_{\{\omega_j\}}\sum_i\left(\log \sum_{j\leq p}\omega_j\phi_j(x_i)\right)-\sum_{j\leq p}\omega_j \end{aligned} \]

We have a similar log-likelihood to the density estimation case.

Under the constraint that \(E(\hat{N}=N)\) we have \(\sum_j\omega_j=n\) and therefore

\[ \hat{\tau} =\operatorname{argmax}_{\{\omega_i\}}\sum_i\left(\log \sum_{j\leq p}\omega_j\phi_j(x_i)\right)-n \]

Note that if we consider the points as a density we find the same is the same as the maximum obtained by considering the points as an inhomogeneous spatial point patter, up to an offset of \(n\), i.e. \(\omega_j\equiv nw_j.\)

Count regression

We can formulate density estimation as a count regression; For “nice” distributions this will be the same as estimating the correct Poisson intensity for every given small region of the state space (e.g. (Gu 1993; Eilers and Marx 1996)). 🏗

Probability over boxes

Consider a box in \(B\subset \mathbb{R}^d\). The probability of any one \(X_i\) falling within that box,

\[P(X_i\subset B)=E\left(\mathbb{I}\{X_i\subset B\}\right)=\int_B dF(x).\]

We know that the expected number of \(X_i\) to fall within that box is \(N\) times the probability of any one falling in that box, i.e.

\[E\left(\sum_{i\leq N}\mathbb{I}\{X_i\subset B\}\right)=N\int_B dF(x)\] and thus

\[P(N(B)=k)=\frac{\exp(-\Lambda(B))\Lambda(B)^k}{k!}.\]

…Where was I going with this? Something to do with linear point process estimation perhaps? 🏗

References