Estimation of the extreme value index and high quantiles under ...

Estimation of the extreme value index and high
quantiles under random censoring
Jan Beirlant (1) & Emmanuel Delafosse (2) & Armelle Guillou (2)
(1) Katholieke Universiteit Leuven, Department of Mathematics, Celestijnenlaan 200B,
3001 Leuven, Belgium
(2) Université Paris VI, L.S.T.A., Boîte 158, 175 rue du Chevaleret, 75013 Paris
Key words and phrases: Pareto index, extreme quantile, censoring, Kaplan-Meier
estimator.
Abstract. In this paper, we consider the estimation problem of the extreme value index
and extreme quantiles in the presence of censoring. Taking into account the fact that our
main motivation is application in insurance, we focus on the Fréchet and Gumbel domains
of attraction. In the case of no-censoring, the most famous estimator of the Pareto index
is the classical Hill estimator (1975). Some adaptations of this estimator in the case of
censoring are proposed and used to build extreme quantile estimators. A theoretical study
of the asymptotic properties of such estimators is started. The finite sample behaviour is
illustrated in a small simulation study and also in a practical insurance example.
Résumé. Dans cet article, nous considérons le problème de l’estimation d’un index
des valeurs extrêmes et de quantiles extrêmes en présence de censure aléatoire. Compte
tenu du fait que notre motivation principale concerne l’application en assurance, nous
nous concentrons sur les domaines d’attraction de Fréchet et de Gumbel. Dans le cas
non censuré, l’estimateur de l’index le plus connu est l’estimateur de Hill (1975). Nous
proposons des adaptations de cet estimateur de l’index dans le cas censuré que nous
utilisons par la suite dans le but d’estimer un quantile extrême. Une étude théorique
des propriétés asymptotiques de ces nouveaux estimateurs est proposée. Par ailleurs, leur
comportement est illustré sur la base de simulations et sur un exemple de données réelles.
Mots-clés: Index de Pareto, quantile extrême, données censurées, estimateur de Kaplan-
Meier.
1. Introduction.
When a data set contains observations within a restricted range of values, but otherwise
not measured, it is called a censored data set. Statistical techniques for analyzing censored
data sets are quite well studied, especially in survival analysis and biostatistics in general
where censoring mechanisms are quite common. Especially the case of right censoring
where some results are known to be at least as large as the reported value, received a
lot of attention. Here we can for instance refer to Cox and Oakes (1984). This then
1

concerns central characteristics of the underlying distribution. The literature on tail or
extreme value analysis for censored data is almost non existing. In Reiss and Thomas
(1997) (section 6.1), Beirlant et al. (1996) (section 2.7) and Beirlant and Guillou (2001) in
case of truncated data, some estimators of tail indices were proposed without any deeper
study on their behaviour. However, important problems such as the estimation of extreme
quantiles apparently were not considered before in general.
Data sets with censored extreme data often occur in insurance when reported payments
cannot be larger than the maximum payment value of the contract. When the reported
payment equals the maximum payment, this real payment can indeed be equal to the
maximum or can be censored. The situation where all data above a fixed value are
censored is referred to as truncation or type I censoring. This case was considered in
Beirlant and Guillou (2001). It can occur when the observations are not the real payments
but the payments as a fraction of the sum insured, in which case the truncation level equals
100%. Here we consider random right censoring. The claim sizes X are possibly censored
by the maximum payment Y . A maximum payment of a given contract is then considered
as a realization of the random variable Y . Different situations can now occur, whether
the censoring values (or maximum payment values) are observed or not.
To be more specific, let X i , i ∈ IN, be independent and identically distributed (i.i.d.)
random variables with common distribution function (df) F and let Y i , i ∈ IN, be a
second i.i.d. sequence with df G. We only observe Z i = X i ∧ Y i , δ i = 1l Xi ≤Y i
, i ∈ IN. We
denote by H the df of Z 1 and let τ H = inf{x : H(x) = 1}, the supremum of the support
of H. We define H 1 (z) = IP(Z > z, δ = 1) = IP(z < X ≤ Y ).
Being motivated by actuarial applications we confine ourselves to the case where sample
maxima from X samples are in the domain of attraction of the Fréchet or Gumbel law.
This typically means that we consider polynomially decreasing tails or exponentially decreasing
tails with infinite right endpoint. We will consequently consider the following
cases:
• Observing (Z, δ), X independent of Y , and both X and Y are in the domain of attraction
of the Fréchet law;
• Observing (Z, δ), X independent of Y , X is in the domain of attraction of the Fréchet
or the Gumbel law, and Y in the domain of attraction of the Fréchet law.
In order to illustrate the methods presented in this paper, we use a liability insurance
example from Frees and Valdez (1998).
2. Estimation techniques.
2.1. Observing (Z, δ), X independent of Y , and both X and Y are in the domain
of attraction of the Fréchet law
2

Supposing that F is of Pareto-type, that is, there exists a positive constant α for which
where l 1 is a slowly varying function at infinity satisfying
1 − F (x) = x −α l 1 (x), (1)
l 1 (λx)
l 1 (x)
→ 1 when x → ∞, for all λ > 0.
In order for the censoring to be not too heavy, it appears natural to assume that the
censoring distribution is also heavy tailed
1 − G(x) = x −β l 2 (x), (2)
for some β > 0 and slowly varying l 2 . Assuming that X and Y are independent, so that
1 − H(x) = (1 − F (x))(1 − G(x)), it now follows that
1 − H(x) = x −(α+β)˜l(x), (3)
with ˜l also a slowly varying function at infinity. These conditions can be restated in terms
of the tail quantile functions as
U F (x) = x 1/α l 1,U (x), U G (x) = x 1/β l 2,U (x), U H (x) = x 1/(α+β)˜lU (x),
with U F (x) = inf{y : F (y) ≥ 1 − 1/x}, x > 1, and l 1,U (x), l 2,U (x) and ˜l U (x) again slowly
varying functions at infinity.
Our goal is( to ) discuss the estimation problem of γ 1 := α −1 and of extremes quantiles
x F,p := U 1
F p with p <
1
. This problem has received a lot of attention in case of nocensoring,
i.e. when X i ≤ Y i for all i = 1, ..., n. The most famous estimator of γ 1 is Hill’s
n
(1975) estimator, given by
H X,k,n = 1 k∑
log X n−i+1,n − log X n−k,n . (4)
k
i=1
Turning to the estimation of high quantiles, the estimator proposed by Weissman (1978)
serves as a reference under Pareto-type models without censoring:
ˆx p,k = X ( k + 1 ) HX,k,n
n−k,n . (5)
(n + 1)p
In case of random right censoring, the likelihood based on E j,t = Z j
, Z t j > t, is changed
into
∏N t ( )
αE
−α−1 δj
( )
j E
−α 1−δj
j ,
j=1
3

leading to the estimator
H (c)
Z,t =
∑ ni=1
log(Z i /t)1l {Zi >t}
∑ ni=1
, (6)
δ i 1l {Zi >t}
while for the extreme quantile estimator we propose to use
ˆx (c)
p,t = t
( 1 − ˆFn (t)
p
) H
(c)
Z,t
, (7)
where ˆF n (x), −∞ < x < τ H denotes the Kaplan-Meier (1958) product limit estimator of
F (x), defined as
1 − ˆF
n∏
[
n (x) = 1 − δ ]
j,n1l Zj,n ≤x
,
n − j + 1
j=1
where Z j,n denote the order statistics associated to Z 1 , ..., Z n and δ j,n := δ k if and only if
Z j,n = Z k .
The corresponding tail probability estimator is now of course given by
IP ˆ
(c) (X > x) = (1 − ˆF n (t)) ( x) (c) −1/H Z,t
. (8)
t
When choosing t = Z n−k,n , we obtain the estimator
H (c)
Z,k,n = ∑ kj=1
(
log(Zn−j+1,n ) − log(Z n−k,n ) )
∑ kj=1
δ n−j+1,n
, (9)
which is the original Hill estimator adapted for right censoring.
We will give also another interpretation for this estimator which is based on a novel
QQ-plot.
2.2. Observing (Z, δ), X independent of Y , X in the domain of attraction of
the Fréchet or Gumbel law, and Y in the domain of attraction of the Fréchet
law
When considering the extension to the case where γ 1 ≥ 0, again as in the no-censoring case
there are mainly two sets of solutions which originated from two different formulations of
the model.
First, the maximum likelihood approach based on POT’s (Peaks over Threshold) is based
on the results given by Balkema and de Haan (1974) and Pickands (1975), stating that
the limit distribution of the absolute exceedances over a threshold t when t → ∞ is given
by a generalized Pareto distribution (GPD). In the case of censoring, we can easily adapt
the likelihood to
k∏ [
fGP D (Ẽj) ] δ j
[
1 − FGP D (Ẽj) ] 1−δ j
j=1
4

where Ẽj = Z j − t if Z j > t and 1 − F GP D (x) = ( )
1 + γ −
1
1x γ 1
. Then, the maximization of
σ
this expression leads to a POT estimator for γ 1 which we further denote by ˆγ t,ML.
(c)
Secondly, we can construct a new estimator based on k upper order statistics for instance
within the framework of the QQ-plot regression technique. For example, in the case of
no-censoring, Beirlant et al. (1996) proposed an estimator of a real-valued index based on
a generalized quantile plot, which takes over the role of the Pareto quantile plot in this
more general setting. More precisely they proposed to look at the graph with coordinates
( n + 1
)
log , log UH j,n , j = 1, ..., n − 1,
j
with UH j,n = X n−j,n H X,j,n . Again this plot becomes ultimately linear for small j with
slope approximating γ 1 . Then, one can construct several regression based estimators, such
as
ˆγ k,UH = 1 k∑
log UH j,n − log UH k+1,n .
k
j=1
From the above it appears natural to define a generalization of ˆγ k,UH to the censoring
case as a slope estimator of the generalized quantile plot adapted for censoring
( (
− log 1 − ˆFn (Z n−j+1,n ) ) , log UH j,n) (c) , (10)
(j = 1, ..., n − 1) where UH (c)
j,n = Z n−j,n H (c)
Z,j,n:
ˆγ (c)
1 ∑ kj=1
k,UH = log UH (c)
k
1
k
j,n − log UH (c)
k+1,n
∑ kj=1
. (11)
δ n−j+1,n
Using one of the abovementioned estimators ˆγ (c)
.,. of γ 1 ≥ 0 we can now propose new
estimators for the quantile x F,p , in the spirit of the one proposed by Dekkers et al. (1989)
in the case of no-censoring:
ˆx (c)
p,t,. = t + ˆγ (c)
.,. t
( 1− ˆFn(t)
)ˆγ (c)
.,.
p
ˆγ (c)
.,.
− 1
. (12)
Under suitable assumptions, we establish the asymptotic properties of our estimators. We
illustrate their behaviour in a small simulation study, but also in a practical insurance
example.
5

Bibliography
[1] Balkema, A. and de Haan, L. (1974). Residual life time at great age, Ann. Probab.,
2, 792-804.
[2] Beirlant, J. and Guillou, A. (2001). Pareto index estimation under moderate right
censoring, Scand. Actuarial J., 2, 111-125.
[3] Beirlant, J. Teugels, J.L. and Vynckier, P. (1996). Practical Analysis of Extreme
Values, Leuven University Press, Leuven.
[4] Beirlant, J., Vynckier, P. and Teugels, J.L. (1996). Excess functions and estimation of
the extreme value index, Bernoulli, 2, 293-318.
[5]Cox, D.R. and Oakes, D (1984). Analysis of Survival Data, Chapman and Hall, New
York.
[6] Dekkers, A.L.M., Einmahl, J.H.J. and de Haan, L. (1989). A moment estimator for
the index of an extreme-value distribution, Ann. Statist. 17, 1833-1855.
[7] Frees, E. and Valdez, E. (1998). Understanding relationships using copulas, North
American Actuarial Journal, 2, 1–15.
[8] Hill, B.M. (1975). A simple general approach to inference about the tail of a distribution,
Ann. Statist., 3, 1163-1174.
[9] Kaplan, E.L. and Meier, P. (1958). Non-parametric estimation from incomplete observations,
J. Amer. Statist. Assoc., 53, 457-481.
[10] Pickands III, J. (1975). Statistical inference using extreme order statistics, Ann.
Statist., 3, 119-131.
[11] Reiss, R.D. and Thomas, M. (1997). Statistical Analysis of Extreme Values with
Applications to Insurance, Finance, Hydrology and Other Fields, Birkhäuser Verlag, Basel.
[12] Weissman, I. (1978). Estimation of parameters and large quantiles based on the k
largest observations. J. Amer. Statist. Assoc. 73, 812-815.
6