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Dependence Modelling in Insurance via Copulas with Skewed Generalised Hyperbolic Marginals

  • Vitali Alexeev , Katja Ignatieva EMAIL logo and Thusitha Liyanage
Published/Copyright: December 20, 2019

Abstract

This paper investigates dependence among insurance claims arising from different lines of business (LoBs). Using bivariate and multivariate portfolios of losses from different LoBs, we analyse the ability of various copulas in conjunction with skewed generalised hyperbolic (GH) marginals to capture the dependence structure between individual insurance risks forming an aggregate risk of the loss portfolio. The general form skewed GH distribution is shown to provide the best fit to univariate loss data. When modelling dependency between LoBs using one-parameter and mixture copula models, we favour models that are capable of generating upper tail dependence, that is, when several LoBs have a strong tendency to exhibit extreme losses simultaneously. We compare the selected models in their ability to quantify risks of multivariate portfolios. By performing an extensive investigation of the in- and out-of-sample Value-at-Risk (VaR) forecasts by analysing VaR exceptions (i.e. observations of realised portfolio value that are greater than the estimated VaR), we demonstrate that the selected models allow to reliably quantify portfolio risk. Our results provide valuable insights with regards to the nature of dependence and fulfils one of the primary objectives of the general insurance providers aiming at assessing total risk of an aggregate portfolio of losses when LoBs are correlated.

JEL Classification: G22; C46; C15

Appendix: copula estimation

Generally, the maximum likelihood technique is applied to estimate parametric copulas. From (1), the density of the random vector X = (X 1, …, Xd ) is given by

(22) f ( x 1 , , x d ; δ 1 , , δ d , θ ) = c { F X 1 ( x 1 ; δ 1 ) , , F X d ( x d ; δ d ) ; θ } j = 1 d f j ( x j ; δ j ) ,

where

(23) c ( u 1 , , u d ) = d C ( u 1 , , u d ) u 1 u d

denotes a copula density. Denoting a vector of parameters α = (δ 1, …, δ d , θ) ∈ ℝ d+1, the likelihood function is given by

(24) L ( α ; x 1 , , x T ) = t = 1 T f ( x 1 , t , , x d , t ; δ 1 , , δ d , θ ) .

Combining (22) and (24), the corresponding log-likelihood function is given by

(25) ( α ; x 1 , , x T ) = t = 1 T ln [ c { F X 1 ( x 1 , t ; δ 1 ) , , F X d ( x d , t ; δ d ) ; θ } ] + t = 1 T j = 1 d ln [ f j ( x j , t ; δ j ) ] .

To maximize this log-likelihood numerically, we perform the inference for marginals (IFM) method, which is a sequential two-step maximum likelihood method, see e.g. McLeish and Small (1988) and Joe (1997). Parameters from the marginals are estimated in the first step as

(26) δ ^ j = arg max δ j ( δ j ) ,

where

(27) j ( δ j ) = t = 1 T ln f j [ x j , t ; δ j ]

denotes the log-likelihood function for each of the marginal distribution j = 1, …, d. Their estimates are then substituted into the copula to obtain the pseudo log-likelihood function

(28) ( θ , δ ^ 1 , , δ ^ d ) = t = 1 T ln [ c { F X 1 ( x 1 , t ; δ ^ 1 ) , , F X d ( x d , t ; δ ^ d ) ; θ } ] ,

which is then maximised with respect to θ to obtain the estimator θ ^ . The estimates α ^ I F M = ( δ 1 ^ , , δ d ^ , θ ^ ) solve the first order condition

(29) ( 1 / δ 1 , , d / δ d , / θ ) = 0.

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Published Online: 2019-12-20

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