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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5.2 Stochastic volatility 73<br />
3. An estimate for the unobserved process θ can be computed from the <strong>state</strong> <strong>space</strong><br />
model given by (5.26) and (5.39) with (5.29) and (5.30) by mak<strong>in</strong>g use <strong>of</strong> the<br />
Kalman smoother as <strong>in</strong>troduced <strong>in</strong> ¢ 3.3.2.<br />
4. Steps #2 and #3 are repeated us<strong>in</strong>g the smoothed estimates <strong>of</strong> θ until convergence<br />
<strong>of</strong> θ, at and bt is achieved.<br />
5. F<strong>in</strong>ally, with the importance density g � θ (i) |y � be<strong>in</strong>g evaluated us<strong>in</strong>g the Kalman<br />
smoother, the likelihood given by (5.38) can be estimated with the samples θ (i)<br />
that are drawn from g � θ (i) |y � via simulation smooth<strong>in</strong>g.<br />
In order to avoid numerical problems, this is usually done by work<strong>in</strong>g with logarithms <strong>of</strong><br />
the likelihood. Let w(θ) = p(y|θ)/g(y|θ) and wi = w � θ (i)� for i = 1, . . . , M. Accord<strong>in</strong>g<br />
to Durb<strong>in</strong> and Koopman (1997) an approximately unbiased estimator <strong>of</strong> log L(ψ) is<br />
given by<br />
log ˆ L(ψ) = log LG(ψ) + log ¯w + s2w , (5.40)<br />
2M ¯w 2<br />
where ¯w = M −1 � M<br />
i=1 wi and s 2 w = (M − 1)−1 � M<br />
i=1 (wi − ¯w) 2 represent the sample<br />
mean and variance <strong>of</strong> wi, respectively. The last term on the right hand side corrects<br />
for the bias that can be shown to arise from tak<strong>in</strong>g logarithms <strong>of</strong> (5.38). Accord<strong>in</strong>g to<br />
Durb<strong>in</strong> and Koopman (2000) the bias is usually so small that it can be neglected <strong>in</strong> practice.<br />
Numerical maximization <strong>of</strong> the simulated loglikelihood (5.40), us<strong>in</strong>g an iterative<br />
numerical optimization method as discussed <strong>in</strong> ¢ 3.4.2, yields the MCL estimate <strong>of</strong> ψ.<br />
Start<strong>in</strong>g values can be obta<strong>in</strong>ed from log LG(y|ψ). The accuracy <strong>of</strong> the approximation<br />
<strong>of</strong> the likelihood function depends positively on M, the number <strong>of</strong> draws. Sandmann<br />
and Koopman (1998) showed that N = 5 is sufficient <strong>in</strong> empirical applications.<br />
In this thesis, the SV model is estimated us<strong>in</strong>g Ox 3.30 by Doornik (2001) together<br />
with the packages SsfPack 2.3 and SsfNonG by Koopman et al. (1999). The relevant<br />
code for implement<strong>in</strong>g the estimation <strong>of</strong> SV <strong>models</strong> <strong>in</strong> Ox has been downloaded from<br />
www.feweb.vu.nl/koopman/sv/.<br />
5.2.3.3 Filter<strong>in</strong>g, smooth<strong>in</strong>g and forecast<strong>in</strong>g<br />
Once estimates for the unknown model parameters have been found as demonstrated<br />
above, another <strong>in</strong>terest focuses on obta<strong>in</strong><strong>in</strong>g filtered, smoothed and predicted estimates<br />
<strong>of</strong> the conditional volatility process. In contrast to the GARCH model, where filtered<br />
estimates <strong>of</strong> the volatility process can be calculated recursively once the parameters<br />
are known, for an SV model the latent volatility has to be estimated. In addition to<br />
the filtered volatility also smoothed estimates are available via the Kalman filter and<br />
smoother (cf. ¢ 3.3); see Sandmann and Koopman (1998) for details.<br />
Accord<strong>in</strong>g to Koopman et al. (2004) the one-step ahead forecast <strong>of</strong> conditional volatility<br />
can be calculated as<br />
ˆσ 2 T +1 = ˆσ2 ∗ exp<br />
�<br />
ˆh T +1|T + 1<br />
2 p �<br />
T +1|T , (5.41)