Abstract
This study proposes a new Markov switching process with clustering effects. In this approach, a hidden Markov chain with a finite number of states modulates the parameters of a self-excited jump process combined to a geometric Brownian motion. Each regime corresponds to a particular economic cycle determining the expected return, the diffusion coefficient and the long-run frequency of clustered jumps. We study first the theoretical properties of this process and we propose a sequential Monte-Carlo method to filter the hidden state variables. We next develop a Markov Chain Monte-Carlo procedure to fit the model to the S&P 500. We find that self-exciting jumps occur mainly during economic recession and nearly disappear in periods of economic growth. Finally, we analyse the impact of such a jump clustering on implied volatilities of European options.
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Notes
Notice that stochastic volatility models also explain assymmetry and high kurtosis. However Cont and Tankov (2004) mention on page 6 that Brownian models with stochastic volatility cannot explain the presence of jumps in price due to the continuity of sample paths.
Notice that if \(C\in [u^{*},\omega _{2})\), the function \(F_{\omega _{1}}(C)\) is equal to
$$\begin{aligned} F_{\omega _{1}}(C):&=-\int _{C}^{\omega _{2}}\frac{du}{-\beta (\omega _{1})-\alpha u+\psi \left( \omega _{1}\,,\,\eta \,u\right) }. \end{aligned}$$We can think to relate thresholds to regimes. For example, if we denote by \(\tilde{\sigma }_{i}^{ML}\) the volatility of the SGBM in the most likely state at time \(t_{i}\), we can assume that \(g(\alpha _{k},i)=\sqrt{\Delta }\tilde{\sigma }_{i}^{ML}\Phi ^{-1}(\alpha _{k})\quad k=1,2\). However, when this method is applied to the S&P 500 data set, fewer jumps are detected during recessions than in periods of growth. This counter intuitive result is explained by the fact that thresholds are proportional to the volatility in each regime. As the volatility is much important during recessions than in other periods, thresholds are also much higher. Consequence: less log-returns exceed thresholds during bad economic times. This observation motivates us to not relate thresholds to regimes and to use instead the standard deviation of the whole sample. This reduces the accuracy of the POT method. However, given that we use it to find an acceptable starting point for the MCMC algorithm, this loss of accuracy has a limited impact on final conclusions.
In theory the acceptance rate can be improved to any desired level. Increasing the number of particles raises considerably the computational time of the estimation procedure.
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Acknowledgements
Donatien Hainaut thanks for its support the Chair “Data Analytics and Models for insurance” of BNP Paribas Cardif, hosted by ISFA (Université Claude Bernard, Lyon) and managed by the “Fondation Du Risque”.
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Appendices
Appendix A: Proofs
Proof of proposition 2.1
As \(\mathcal {F}_{0}\subset \mathcal {F}_{0}\vee \mathcal {G}_{t}\), the expectation of \(\lambda _{t}\) is nested as follows:
and if we remember the expression (9) for the intensity, we infer that
Let us denote by \(\chi (t,l)\), the random measure of \(L_{t}|\mathcal {F}_{0}\vee \mathcal {G}_{t}\). \(\chi (t,l)\) is such that \(L_{t}|\mathcal {F}_{0}\vee \mathcal {G}_{t}=\int _{0}^{t}\int _{0}^{\infty }\chi (ds,dl)\) and \(d\left( L_{t}|\mathcal {F}_{0}\vee \mathcal {G}_{t}\right) =\int _{0}^{\infty }\chi (ds,dl)\). Given that the integrand \(\eta e^{\alpha (s-t)}\) is time integrable on \(\mathbb {R}^{+}\), independent from the random measure \(\chi \), we infer that
According to the Itô’s lemma for semi-martingale and by construction of \(L_{s}\), \(L_{s}\) is solution of the following SDE:
Given that the jump J is independent from \(N_{t}\), the expectation of \(dL_{s}\) with respect to \(\mathcal {F}_{0}\vee \mathcal {G}_{t}\) is then
for \(s\le t\). Furthermore, conditionally to the filtration of \(\lambda _{t}\) denoted \(\left( \mathcal {L}_{t}\right) _{t\ge 0}\), \(\left( N_{t}\right) _{t\ge 0}\) is a Poisson random variable of intensity and expectation equal to \(\int _{0}^{t-}\lambda _{u}du\). Using nested expectations allows us to infer that
Combining (33) and (34) leads then to
and
If we derive this last expression with respect to time, we find that \(\mathbb {E}\left( \lambda _{t}|\mathcal {F}_{0}\vee \mathcal {G}_{t}\right) \) is solution of an ordinary differential equation (ODE):
The solution of this ODE is the following function:
and the \(\mathcal {F}_{0}\) conditional expectation is given by:
The expected level of mean reversion at time s is the sum of \(\bar{\theta }_{j}\) for \(j=1\) to N weighted by the probabilities of transition:
and if I is the \(N\times N\) identity matrix, the integral in (36) may be rewritten as follows:
\(\square \)
Proof of Corollary 2.2
Conditionally to \(\mathcal {G}_{t}\), this expectation is equal to
We infer from this relation that the derivative with respect to time is given by
and we can conclude that \(\mathbb {E}\left( \theta _{t}\lambda _{t}|\mathcal {F}_{0}\right) \) is well given by Eq. (12). \(\square \)
Proof of proposition 2.3
Let us introduce the following notations: \(f(t,N_{t},\lambda _{t},\delta _{t})=\lambda _{t}^{2}\), according to the Itô’s lemma for semi-martingale, the infinitesimal generator of this function is given by:
On the other hand, we can prove that
The derivative of this expectation with respect to time is equal to its expected infinitesimal generator:
that allows us to infer Eq. (13). It is easy to check that the right term of Eq. (38) is linear and Lipschitz. Then, according to the theorem of Cauchy-Lipschitz the solution exists and is unique. \(\square \)
Proof of proposition 2.4
If we note \(f(t,N_{t},\lambda _{t},\delta _{t})=\mathbb {E}\left( \omega ^{N_{s}}\,|\,\mathcal {F}_{t}\right) \), f is solution of an Itô’s equation for semi martingale. If \(\delta _{t}=e_{i}\) then:
If we assume that f is an exponential affine function of \(\lambda _{t}\) and \(N_{t}\):
where \(A(t,s,e_{i})\) for \(i=1\) to N, B(t, s), C(t, s) are time dependent functions, the partial derivatives of f are given by:
The integral in Eq. (39) is rewritten as follows:
As \(q_{ii}=-\sum _{i\ne j}^{N}q_{i,j}\), the last term of Eq. (39) becomes
Then
from which we deduce that \(C(t,s)=\ln \omega \). Regrouping terms allows to infer that
or that
If we define \(\tilde{A}(t,s,e_{i})=e^{A(t,s,e_{i})}\), the first equations can finally be put in matrix form as:
\(\square \)
Proof of proposition 2.5
If we note \(f(t,X_{t},\lambda _{t},\delta _{t})=\mathbb {E}\left( e^{\omega _{1}X_{s}+\omega _{2}\lambda _{s}}\,|\,\mathcal {F}_{t}\right) \), f is solution of an Itô’s equation for semi-martingale. If \(\delta _{t}=e_{i}\) then:
If we assume that f is an exponential affine function of \(\lambda _{t}\) and \(X_{t}\):
where \(A(t,s,e_{i})\) for \(i=1\) to N, B(t, s), C(t, s) are time dependent functions, the partial derivatives of f are given by:
and the integrand in Eq. (41) is rewritten as follows:
As \(q_{ii}=-\sum _{i\ne j}^{N}q_{i,j}\), the last term of the Itô equation is also equal to
injecting these expressions into Eq. (41), leads to the following relation:
from which we deduce that \(C(t,s)=\omega _{1}\). Regrouping terms allows to infer that A and B are solutions of a system of ODE’s:
From this last relation, we deduce that
If we define \(\tilde{A}(t,s)=(e^{A(t,s,e_{i})})_{i=1,\ldots ,N}\), the first equations can finally be put in matrix form as:
\(\square \)
Proof of proposition
From the previous results, we know that B(t, s) is solution of the ODE:
with the terminal condition: \(B(s,s)=\omega _{2}\). If we set \(B(t,s)=C(s-t)\) and \(\tau =s-t\). Then \(\frac{\partial }{\partial t}B=\frac{\partial }{\partial \tau }C\frac{\partial \tau }{\partial t}=-\frac{\partial }{\partial \tau }C\) and
The left hand side is denoted h(C). Due to the convexity of \(\psi \left( .\right) \) there is only one point \(u^{*}\) such that \(h(u)=0\). This equation is indeed equivalent to
We rewrite the Eq. (43) as follows,
As \(C(0)=\omega _{2}\), by direct integration, we have that
with \(C\in [\omega _{2},u^{*})\) or \(C\in [u^{*},\omega _{2})\). Remark that if \(C=u^{*}\) then \(\tau =+\infty \) as the numerator converges to zero. If we define the function on the left hand side as
then \(F_{\omega _{1}}(C)=\tau \) and \(C=F_{\omega _{1}}^{-1}(\tau )\) or \(B(t)=F_{\omega _{1}}^{-1}(s-t)\). \(\square \)
Appendix B: Particle filter
The structure of the particle filter algorithm is the following:
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1.
Initial step: draw M values of \(v_{0}^{(i)}\) for \(i=1,\ldots ,M\), from an initial distribution \(p(v_{0})\)
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2.
For \(j=1\,:\,T\)
Prediction step: draw a sample of \(\Delta L_{j}^{(i)}\) and \(\delta _{j}^{(i)}\) and update \(\lambda _{j}^{(i)}\), \(\mu _{j}^{b\,(i)}\)\(\sigma _{j}^{(i)}\) , \(\theta _{j}^{(i)}\) using the relations
$$\begin{aligned}&\lambda _{j}^{(i)}=\lambda _{j-1}^{(i)}+\alpha (\theta _{j-1}^{(i)}-\lambda _{j-1}^{(i)})\Delta +\eta {\varvec{\Delta }}L_{j}^{(i)}\\&\mu _{j}^{(i)}=\delta _{j}^{(i)}\bar{\mu }\quad \sigma _{j}^{(i)}=\delta _{j}^{(i)}\bar{\sigma }\quad \theta _{j}^{(i)}=\delta _{j}^{(i)}\bar{\theta } \end{aligned}$$Correction step: the particle \(v_{j}^{(i)}\) has a probability of occurrence equal to \(w_{j}^{(i)}=\frac{p(x_{j}\,|\,v_{j})}{\sum _{i=1:M}p(x_{j}\,|\,v_{j})}\) where
$$\begin{aligned} p(x_{j}\,|\,v_{j}^{(i)})&\sim \mathcal {N}\left( \left( \mu _{j}^{(i)}-\frac{\sigma _{j}^{(i)2}}{2}-\lambda _{j}^{(i)}\mathbb {E}\left( e^{J}-1\right) \right) \Delta -{\varvec{\Delta }}L_{j}^{'(i)}\,,\,\sigma _{j}^{(i)}\sqrt{\Delta }\right) \end{aligned}$$Resampling step: resample with replacement M particles according to the importance weights \(w_{j}^{(i)}\). The new importance weights are set to \(w_{j}^{(i)}=\frac{1}{M}\).
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Hainaut, D., Moraux, F. A switching self-exciting jump diffusion process for stock prices. Ann Finance 15, 267–306 (2019). https://doi.org/10.1007/s10436-018-0340-5
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DOI: https://doi.org/10.1007/s10436-018-0340-5