The *ReIns* package contains functions from the book
“Reinsurance: Actuarial and Statistical Aspects” (2017) by Hansjörg
Albrecher, Jan Beirlant and Jef Teugels.

It contains implementations of

Basic extreme value theory (EVT) estimators and graphical methods as described in “Statistics of Extremes: Theory and Applications” (2004) of Jan Beirlant, Yuri Goegebeur, Johan Segers and Jef Teugels.

EVT estimators and graphical methods adapted for censored and/or truncated data.

Splicing of mixed Erlang distributions with EVT distributions (Pareto, GPD).

Value-at-Risk (VaR), Conditional Tail Expectation (CTE) and excess-loss premium estimates.

This vignette describes how to use the most important functions of
the package. We split this into several sections: *datasets, graphical
methods, estimators of the extreme
value index, estimators of
quantiles and return periods, censored
data, global fits using
splicing, risk measures, distributions* and *approximations of
the distribution function*.

Three datasets are available: **Norwegian fire insurance
data** (`norwegianfire`

), **SOA group medical
insurance data** (`soa`

) and **Secura Re
automobile reinsurance data** (`secura`

). These
datasets were already discussed in Beirlant et
al. (2004).

The illustrations will be done using the Norwegian Fire insurance dataset which contains fire insurance claims for a Norwegian insurance company for the period 1972 to 1992. The sizes of the fire insurance claims are expressed in 1000 NOK.

QQ-plots and their derivative plots are an essential part of extreme value theory. We focus on four important types: the exponential QQ-plot, Pareto QQ-plot, log-normal QQ-plot and Weibull QQ-plot, see Section 4.1 in Albrecher et al. (2017).

The **exponential QQ-plot** can be easily drawn using
`ExpQQ`

. Its derivative plot, also dubbed **mean excess
plot**, is key in determining what type of distribution the data
comes from. The mean excess values \(e_{k,n}\) are plotted using
`MeanExcess`

and one has the choice to plot them versus the
order statistics \(X_{n-k,n}\)
(`k=FALSE`

) or versus the number of exceedances \(k\) (`k=TRUE`

).

In this case we see that the mean excess plot is more or less linearly increasing as a function of \(X_{n-k,n}\) which indicates that the data may come from a Pareto distribution. The exponential QQ-plot is not linear at all but concave which reinforces the conclusion drawn from the mean excess plot.

Taking the logarithm of Pareto distributed data gives exponentially
distributed data, so the **Pareto QQ-plot** and the
exponential QQ-plot are closely related. Using the commands
`ParetoQQ`

and `ParetoQQ_der`

, the Pareto QQ-plot
and its derivative plot can be drawn. Note that these derivatives are
nothing more than the Hill estimates.

The Pareto QQ-plot is now linear which indicates that the Pareto distribution is suitable. The derivative plots can be used to estimate the tail index \(\gamma=1/\alpha\) of the Pareto distribution (cf. Hill estimator).

One can also consider the **log-normal QQ-plot**
(`LognormalQQ`

) and its derivative plot
(`LognormalQQ_der`

). It is clear that a log-normal
distribution is not suitable for this data.

Finally, we look at the **Weibull QQ-plot**
(`WeibullQQ`

) and its derivative plot
(`WeibullQQ_der`

). The concave shape indicates that the
Weibull distribution is not suitable for this data.

More details on the estimators described in this section can be found in Section 4.2 in Albrecher et al. (2017).

The most famous estimator for the EVI \(\gamma\) is the **Hill
estimator** which can be obtained by fitting the (strict) Pareto
distribution to the relative excesses \(X/X_{n-k,n}\) using Maximum Likelihood
Estimation (MLE). The typical Hill plot can be made using
`Hill`

.

The bias of the Hill estimator can be problematic, hence one can
consider a **bias-reduced estimator** which uses a
regression-type approach (`Hill.2oQV`

). Another solution is
to use the **EPD estimator** (`EPD`

) which fits
the extended Pareto distribution instead of the ordinary Pareto
distribution to the relative excesses.

Using these bias-reduced estimators can give an idea about a good choice for \(k\). Suitable choices of \(k\) are values where the two plots intersect and which are not too low (otherwise the variance is too high). This yields an estimate for \(\gamma\) around 0.75 (and \(k\) around 3500).

The previous estimators can only be used when \(\gamma\) is strictly positive. Therefore,
the **generalised QQ-plot** was proposed. This QQ-plot is a
generalisation of the Pareto QQ-plot and can also have negative, or
zero, slopes. The function `genQQ`

needs the Hill estimates
as input.

We see that the generalised QQ-plot is strictly increasing which indicates a strictly positive \(\gamma\).

The **generalised Hill estimator**
(`genHill`

) is an estimator for the slope of the generalised
QQ-plot. Alternatively, one can also consider the **moment
estimator** (`Moment`

) and the
**Peaks-Over-Threshold estimator** (`GPDmle`

)
which fits the Generalised Pareto Distribution (GPD) to the excesses
\(X-X_{n-k,n}\) using Maximum
Likelihood Estimation (MLE). The Peaks-Over-Threshold estimator can
however be very time consuming on large datasets! Therefore, it is
omitted in this example.

The generalised Hill estimator and the moment estimator indicate that values for \(\gamma\) around 0.75 are indeed suitable.

The previously discussed estimators can be used to estimate large quantiles or small exceedance probabilities and corresponding high return periods. These estimators are also described in Section 4.2 in Albrecher et al. (2017).

We can for example estimate the 99.5% quantile using Hill estimates
(`Quant`

) or using generalised Hill estimates
(`QuantGH`

), GPD estimates (`QuantGPD`

) and moment
estimates (`QuantMOM`

).

All three estimators (we exclude the GPD estimator again) suggest a 99.5% VaR value around 35 000 000 NOK.

Estimating the return period of the value 100 000 000 NOK using the
same three estimators (using `Return`

, `ReturnGH`

and `ReturnMOM`

) gives an estimate around 800 (claims).

Several EVT estimators for right censored data are included:

`cExpQQ`

,`cLognormalQQ`

,`cParetoQQ`

,`cWeibullQQ`

: QQ-plots adapted for right censoring.`cHill`

,`cEPD`

,`cgenHill`

,`cGPD`

,`cMoment`

: estimators for the EVI.`cQuant`

,`cQuantGH`

,`cQuantGPD`

,`cQuantMOM`

: estimators for extreme quantiles.`cProb`

,`cProbEPD`

,`cProbGH`

,`cProbGPD`

,`cProbMOM`

: estimators for small exceedence probabilities.

In the example below we plot estimates for the EVI for a simulated sample of right censored data.

We also implemented EVT functions for the more general case of interval censored data:

`MeanExcess_TB`

: mean excess plot adapted for interval censoring (using the Turnbull estimator for the survival function).`icParetoQQ`

: Pareto QQ-plot adapted for interval censoring.`icHill`

: estimator for (positive) EVI.

In the example below we make the mean excess plot for a simulated sample of right censored data.

```
# Set seed
set.seed(29072016)
# Pareto random sample
X <- rpareto(500, shape=2)
# Censoring variable
Y <- rpareto(500, shape=1)
# Observed sample
Z <- pmin(X, Y)
# Censoring indicator
censored <- (X>Y)
# Right boundary
U <- Z
U[censored] <- Inf
# Mean excess plot
MeanExcess_TB(Z, U, censored, k=FALSE)
```

The previous sections dealt with fitting a suitable distribution for
the tail of the data. One usually wants a fit for the whole
distribution. We therefore propose the **splicing** of a
Mixed Erlang (ME) distribution for the body and an extreme value
distribution, i.e. Pareto or GPD, for the tail. See Section 4.3 in
Albrecher et al. (2017) for more details. We consider three possible
fitting procedures:

`SpliceFitPareto`

: fits a splicing model with a ME distribution and Pareto distribution(s) to possibly truncated data.

`SpliceFitGPD`

: fits a splicing model with a ME distribution and a GPD to possibly lower truncated data. This procedure cannot handle upper truncation.`SpliceFiticPareto`

: fits a splicing model with a ME distribution and a Pareto distribution to possibly censored and/or truncated data.

At first, one has to determine (a) suitable splicing point(s). We do this using the mean excess plot. Linear upward pieces indicate that the Pareto distribution is suitable, linear downward pieces suggest a truncated Pareto distribution.

```
# Mean excess plot
MeanExcess(size)
# Add vertical line at 50% and 99.6% quantiles of the data
abline(v=quantile(size, c(0.5,0.996)), lty=2)
```

For the Norwegian fire insurance data, we choose splicing points at the 50% and 99.6% quantiles. This means that the spliced distribution is a ME distribution between 0 and the first splicing point, a (truncated) Pareto distribution between the first and second splicing point and another Pareto distribution after the second splicing point.

Using `SpliceFitPareto`

we can fit this spliced
distribution. This can take some time so we create a
`SpliceFit`

object with the obtained parameters to use in the
remainder. A `summary`

method is available for these objects
which summarises the spliced distribution.

```
# Splicing of Mixed Erlang (ME) and 2 Pareto pieces
# Use 3 as initial value for M
spliceFit <- SpliceFitPareto(size, const=c(0.5,0.996), M=3)
```

```
# Create MEfit object
mefit <- MEfit(p=1, shape=17, theta=44.28, M=1)
# Create EVTfit object
evtfit <- EVTfit(gamma=c(0.80,0.66), endpoint=c(37930,Inf))
# Create SpliceFit object
splicefit <- SpliceFit(const=c(0.5,0.996), trunclower=0, t=c(1020,37930), type=c("ME","TPa","Pa"),
MEfit=mefit, EVTfit=evtfit)
# Show summary
summary(splicefit)
```

```
##
## ----------------------------------------
## Summary of splicing fit
## ----------------------------------------
##
## const = (0.5, 0.996)
##
## pi = (0.5, 0.496, 0.004)
##
## t0 = 0
##
## t = (1 020, 37 930)
##
## type = (ME, TPa, Pa)
##
## * * * * * * * * * * * * * * * * * * * *
##
## p = 1
##
## r = 17
##
## theta = 44.28
##
## M = 1
##
## * * * * * * * * * * * * * * * * * * * *
##
## gamma = (0.8, 0.66)
##
## endpoint = (37 930, Inf)
```

To see how well the spliced distribution fits the data, we use three tools:

`SpliceECDF`

: plot of the fitted survival function and the empirical survival function with confidence bounds.

`SplicePP`

: plot of the fitted survival function vs. the empirical survival function. The plot with minus-log scales is most informative for the tails.`SpliceQQ`

: plot of the fitted quantile function vs. the empirical quantile function.

Similar functions for censored data are implemented as well:
`SpliceTB`

, `SplicePP_TB`

and
`SpliceQQ_TB`

.

We see that the fitted spliced distribution approximates the empirical distribution quite well.

Using the fitted spliced distribution, one can compute estimates for excess-loss premiums, Value-at-Risk (VaR) and Conditional Tail Expectation (CTE). Moreover, data can be simulated from the spliced distribution. See Section 4.6 in Albrecher et al. (2017) for more details.

A commonly used risk measure is the **Value-at-Risk**
(`VaR`

). This is nothing more than the quantile \(Q(1-p)\) of a distribution.

In recent years, the **Conditional Tail Expectation**
(`CTE`

) is gaining importance. It is the conditional
expectation of the data above a certain VaR. It is therefore always a
larger than the corresponding VaR. When the CDF is continuous, which is
the case for the ME-Pa and ME-GPD splicing models, the CTE is equal to
the Tail Value-at-Risk (TVaR).

Simulation of losses is useful for aggregate loss calculations,
e.g. when losses are not independent, and to determine risk measures.
The function `rSplice`

simulates data from the fitted spliced
distribution.

Several distributions have been implemented in the package:

- The Burr distribution (type XII):
`burr`

. - The Extended Pareto Distribution (EPD):
`epd`

. - The Fréchet distribution:
`frechet`

. - The Generalised Pareto Distribution (GPD):
`gpd`

. - The Pareto distribution:
`pareto`

. - The spliced distribution consisting of a Mixed Erlang distribution
and a Pareto or GP distribution:
`splice`

.

Moreover, several upper truncated distributions are included:

- The truncated Burr distribution:
`tburr`

. - The truncated exponential distribution:
`texp`

. - The truncated Fréchet distribution:
`tfrechet`

. - The truncated GPD:
`tgpd`

. - The truncated log-normal distribution:
`tlnorm`

. - The truncated Pareto distribution:
`tpareto`

. - The truncated Weibull distribution:
`tweibull`

.

It is often very useful to approximate a distribution using the first moments. See Section 6.2 in Albrecher et al. (2017) for more details on the approximations discussed in this section.

Several classical approximations are implemented in the function
`pClas`

:

The

**normal approximation**(`method="normal"`

) for the CDF of the r.v. \(X\) is defined as \[F_X(x) \approx \Phi((x-\mu)/\sigma)\] where \(\mu\) and \(\sigma^2\) are the mean and variance of \(X\), respectively, and \(\Phi\) is the CDF of the standard normal distribution.This approximation can be improved when the skewness parameter

\[\nu=E((X-\mu)^3/\sigma^3)\] is available. The**normal-power approximation**(`method="normal-power"`

) of the CDF is then given by \[F_X(x) \approx \Phi\left(\sqrt{9/\nu^2 + 6z/\nu+1}-3/\nu\right)\] for \(z=(x-\mu)/\sigma\geq 1\) and \(9/\nu^2 + 6z/\nu+1\geq 0\).The

**shifted Gamma approximation**(`method="shifted Gamma"`

) uses the approximation \[X \approx \Gamma\left(4/\nu^2, 2/(\nu\times\sigma)\right) + \mu -2\sigma/\nu.\] Here, we need again that \(\nu>0\).The

**normal approximation to the shifted Gamma distribution**(`method="shifted Gamma normal"`

) approximates the CDF of \(X\) as \[F_X(x) \approx \Phi\left(\sqrt{16/\nu^2 + 8z/\nu}-\sqrt{16/\nu^2-1}\right)\] for \(z=(x-\mu)/\sigma\geq 1\). We need again that \(\nu>0\).

These methods need the mean, variance and possibly the skewness coefficient (all four methods except the normal approximation) as input. Below, the distribution function of the chi-squared distribution is approximated using the four methods.

```
# Chi-squared sample
X <- rchisq(1000, 2)
x <- seq(0,10,0.01)
# Classical approximations
p1 <- pClas(x, mean(X), var(X))
p2 <- pClas(x, mean(X), var(X), mean((X-mean(X))^3)/sd(X)^3, method="normal-power")
p3 <- pClas(x, mean(X), var(X), mean((X-mean(X))^3)/sd(X)^3, method="shifted Gamma")
p4 <- pClas(x, mean(X), var(X), mean((X-mean(X))^3)/sd(X)^3, method="shifted Gamma normal")
# True probabilities
p <- pchisq(x, 2)
# Plot true and estimated probabilities
plot(x, p, type="l", ylab="F(x)", ylim=c(0,1), col="red")
lines(x, p1, lty=2)
lines(x, p2, lty=3, col="green")
lines(x, p3, lty=4)
lines(x, p4, lty=5, col="blue")
legend("bottomright", c("True CDF", "normal approximation", "normal-power approximation",
"shifted Gamma approximation", "shifted Gamma normal approximation"),
lty=1:5, col=c("red", "black", "green", "black", "blue"), lwd=2)
```

Based on the theory of orthogonal polynomial expansions, the normal
approximation can be improved. The **Gram-Charlier
approximation** (`pGC`

) is an improvement of the
normal approximation using Hermite polynomials. Another commonly used
approximation is the **Edgeworth expansion**
(`pEdge`

).

Denote the standard normal PDF and CDF respectively by \(\phi\) and \(\Phi\). Let \(\mu\) be the first moment, \(\sigma^2=E((X-\mu)^2)\) the variance, \(\mu_3=E((X-\mu)^3)\) the third central moment and \(\mu_4=E((X-\mu)^4)\) the fourth central moment of the random variable \(X\). The corresponding cumulants are given by \(\kappa_1=\mu\), \(\kappa_2=\sigma^2\), \(\kappa_3=\mu_3\) and \(\kappa_4=\mu_4-3\sigma^4\). Now consider the random variable \(Z=(X-\mu)/\sigma\), which has cumulants 0, 1, \(\nu=\kappa_3/\sigma^3\) and \(k=\kappa_4/\sigma^4=\mu_4/\sigma^4-3\).

The **Gram-Charlier approximation** (`pGC`

)
for the CDF of \(X\) is given by \[\hat{F}_{GC}(x) = \Phi(z) + \phi(z) (-\nu/6
h_2(z)- k/24h_3(z))\] with \(h_2(z)=z^2-1\), \(h_3(z)=z^3-3z\) and \(z=(x-\mu)/\sigma\).

The **Edgeworth expansion** (`pEdge`

) for the
CDF of \(X\) is given by \[\hat{F}_{E}(x) = \Phi(z) + \phi(z) (-\nu/6
h_2(z)- (3k \times h_3(z)+\nu^2h_5(z))/72)\] with \(h_2(z)=z^2-1\), \(h_3(z)=z^3-3z\), \(h_5(z)=z^5-10z^3+15z\) and \(z=(x-\mu)/\sigma\).

Both approximations need the first four raw moments as input when
`raw=TRUE`

or otherwise the mean, variance, skewness
coefficient and kurtosis. Applying the two approximations to the same
chi-squared sample as before gives following plot.

```
x <- seq(0, 10, 0.01)
# Empirical moments
moments = c(mean(X), mean(X^2), mean(X^3), mean(X^4))
# Gram-Charlier approximation
p1 <- pGC(x, moments)
# Edgeworth approximation
p2 <- pEdge(x, moments)
# Normal approximation
p3 <- pClas(x, mean(X), var(X))
# True probabilities
p <- pchisq(x, 2)
# Plot true and estimated probabilities
plot(x, p, type="l", ylab="F(x)", ylim=c(0,1), col="red")
lines(x, p1, lty=2)
lines(x, p2, lty=3)
lines(x, p3, lty=4, col="blue")
legend("bottomright", c("True CDF", "GC approximation",
"Edgeworth approximation", "Normal approximation"),
col=c("red", "black", "black", "blue"), lty=1:4, lwd=2)
```

Albrecher, Hansjörg, Jan Beirlant, and Jef Teugels. 2017.
*Reinsurance: Actuarial and Statistical Aspects*. Wiley,
Chichester.

Beirlant, Jan, Yuri Goegebeur, Johan Segers, and Jef Teugels. 2004.
*Statistics of Extremes: Theory and Applications*. Wiley,
Chichester.