Applications of state space models in finance
Applications of state space models in finance
Applications of state space models in finance
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28 3 L<strong>in</strong>ear Gaussian <strong>state</strong> <strong>space</strong> <strong>models</strong> and the Kalman filter<br />
observations, y 1, . . . , y T , which are assumed to be IID, the jo<strong>in</strong>t density is given by the<br />
product <strong>of</strong> the <strong>in</strong>dividual densities, denoted by p(·):<br />
L(y, ψ) = p(y 1, . . . , y T ) =<br />
T�<br />
p(yt), (3.52)<br />
where L(y, ψ) is the jo<strong>in</strong>t probability density function <strong>of</strong> the t-th set <strong>of</strong> observations.<br />
When the jo<strong>in</strong>t density is evaluated at a given data set, L(y, ψ) is referred to as a<br />
likelihood function <strong>of</strong> the model. To avoid computational difficulties caused by extreme<br />
numbers that may result from the product above, it is generally simpler to work with<br />
the natural logarithm <strong>of</strong> the likelihood function:<br />
log L(y, ψ) =<br />
t=1<br />
T�<br />
log p(yt ). (3.53)<br />
The likelihood function and its logarithm are <strong>of</strong>ten simply denoted as L(y) and log L(y),<br />
respectively. If the vector <strong>of</strong> parameters is identifiable, the ML estimate <strong>of</strong> the parameters<br />
( ˆ ψ) is found by maximiz<strong>in</strong>g the likelihood with respect to ψ. 7 For a general <strong>in</strong>troduction<br />
<strong>in</strong>to the methodology <strong>of</strong> ML, see, for example, Greene (2003, ¢ 17) or Davidson<br />
and MacK<strong>in</strong>non (2004, ¢ 10).<br />
3.4.1.1 Prediction error decomposition<br />
As the observations for time series <strong>models</strong> are not generally <strong>in</strong>dependent, Equation (3.53)<br />
is replaced by a probability density function. The distribution <strong>of</strong> y t is conditioned on<br />
Y t−1, the <strong>in</strong>formation set at time t − 1:<br />
log L(y) =<br />
T�<br />
t=1<br />
t=1<br />
log p(y t |Y t−1), (3.54)<br />
with Y t = {y 1, . . . , y t} and p(y 1|Y 0) := p(y 1).<br />
If the observation disturbances and the <strong>in</strong>itial <strong>state</strong> vector <strong>in</strong> the general <strong>state</strong> <strong>space</strong><br />
model (3.1)–(3.7) have a multivariate normal distribution, it can be shown that the<br />
conditional distribution <strong>of</strong> y t itself is normal with conditional mean<br />
and conditional covariance<br />
E(y t|Y t−1) = Ztat, (3.55)<br />
V ar(y t |Y t−1) = F t. (3.56)<br />
The variance <strong>of</strong> the one-step ahead forecast error, F t, is def<strong>in</strong>ed as <strong>in</strong> (3.16). For Gaussian<br />
<strong>models</strong>, y t is conditionally distributed as N(Ztat, F t) with conditional probability<br />
density function<br />
p(yt|Y t−1) = 1<br />
2π |F t| 1<br />
�<br />
2 exp − 1<br />
2 v′ tF −1<br />
�<br />
t vt . (3.57)<br />
7 Identifiability means that the estimation yields a unique determ<strong>in</strong>ation <strong>of</strong> the parameter<br />
estimates for a given set <strong>of</strong> data. For more details on the identifiability <strong>of</strong> structural time series<br />
<strong>models</strong>, see Harvey (1989, 4.4).