A new weighted Weibull distribution has been defined and studied. Some mathematical properties of the distribution have been studied and the method of maximum likelihood was proposed for estimating the parameters of the distribution. The usefulness of the new distribution was demonstrated by applying it to a real lifetime dataset

Nasiru, Suleman

English

European Scientific Journal, ESJ

European Scientific Journal March  2015 edition vol.11, No.9 ISSN: 1857 – 7881 (Print)  e - ISSN 1857- 7431 134 ANOTHER WEIGHTED WEIBULL DISTRIBUTION FROM AZZALINI’S FAMILY    Suleman Nasiru, MSc. Department of Statistics, Faculty of Mathematical Sciences, University for Development Studies, Navrongo, Upper East Region, Ghana, West Africa   Abstract   A new weighted Weibull distribution has been defined and studied. Some mathematical properties of the distribution have been studied and the method of maximum likelihood was proposed for estimating the parameters of the distribution. The usefulness of the new distribution was demonstrated by applying it to a real lifetime dataset.  Keywords: Weighted Weibull, Order statistic, Maximum likelihood estimation,   Azzalini  Introduction  The Weibull distribution has received much attention in literature because of the advantage it has over other distributions in modeling lifetime data. However, researchers continue to develop different generalizations of the Weibull distribution to increase its flexibility in modeling lifetime data. Merovci and Elbatal (2015) developed the Weibull-Rayleigh distribution and demonstrated its application using lifetime data. Almalki and Yuan (2012) presented the new modified Weibull distribution by combining the Weibull and the modified Weibull distribution in a serial system. The hazard function of the newly proposed distribution is the sum of the Weibull hazard function and a modified Weibull hazard function.  Another generalization of the Weibull distribution is the exponentiated Weibull distribution of Mudholkar and Srivastava (1993). Mudholkar et al. (1995) and Mudholkar and Huston (1996) further studied the exponentiated Weibull distribution with some application to bus-motor failure data and flood. Pal et al. (2006) gave a re-introduction of the exponentiated Weibull distribution in more details. Al-Saleh and Agarwal (2006) proposed another extended version of the Weibull distribution. They demonstrated that the hazard function can exhibit unimodal and bathtub shapes. Xie and Lai (1996) developed the additive Weibull distribution with bathtub shaped hazard function obtained as the sum of two hazard functions European Scientific Journal March  2015 edition vol.11, No.9 ISSN: 1857 – 7881 (Print)  e - ISSN 1857- 7431 135 of the Weibull distribution. Zhang and Xie (2007) employed the Marshall and Olkin (1997) approach of adding new parameter to a distribution to propose the extended Weibull distribution. Lai et al. (2003) proposed a modification of the Weibull distribution by multiplying the Weibull cumulative hazard function by 𝑒𝜆𝑥 and studied its properties.  In this article, a new generalization of the Weibull distribution based on a modified weighted version of Azzalini’s (1985) approach has been proposed. If 𝑔0(𝑥) is a probability density function (pdf) and ?̅?0(𝑥) is the corresponding survival function such that the cumulative distribution function (cdf),𝐺0(𝑥), exist; Then the new weighted distribution is defined as: 𝑓(𝑥;  𝛼, 𝜃, 𝜆) = 𝐾𝑔0(𝑥)?̅?0(𝜆𝑥)                                                   (1) where 𝐾 is a normalizing constant.  Weighted Weibull Distribution  In this section, the density of the weighted Weibull distribution has been derived based on the definition given in equation (1). Consider a two parameter Weibull distribution with pdf given by: 𝑔0(𝑥) = 𝛼𝜃𝑥𝜃−1𝑒(−𝛼𝑥𝜃),      𝑥 > 0, 𝛼 > 0, 𝜃 > 0              (2)  The cdf is given by: 𝐺0(𝑥) = 1 − 𝑒(−𝛼𝑥𝜃)                                                                   (3)  The survival function is given by: ?̅?0(𝑥) = 𝑒(−𝛼𝑥𝜃)                                                                             (4)   Using equations (1), (2) and (4) the pdf of the weighted weibull distribution is defined as: 𝑓(𝑥;  𝛼, 𝜃, 𝜆) = (1 + 𝜆𝜃)𝛼𝜃𝑥𝜃−1𝑒−(𝛼𝑥𝜃+𝛼(𝜆𝑥)𝜃)                       (5)  The corresponding cdf  of the weighted Weibull distribution is given by: 𝐹(𝑥;  𝛼, 𝜃, 𝜆) = 1 − 𝑒−(𝛼𝑥𝜃+𝛼(𝜆𝑥)𝜃)                                             (6) where 𝛼 is a scale parameter and,  𝜃 and  𝜆  are shape parameters. Figure 1 and 2 illustrates possible shapes of the pdf and the cdf of the weighted Weibull distribution for some selected values of the parameters 𝛼, 𝜃  and 𝜆. European Scientific Journal March  2015 edition vol.11, No.9 ISSN: 1857 – 7881 (Print)  e - ISSN 1857- 7431 136  Figure 1: Probability density function of weighted Weibull distribution   Figure 2: Cumulative distribution function of weighted Weibull distribution   The survival function is given by: ?̅?(𝑥;  𝛼, 𝜃, 𝜆) = 1 − 𝐹(𝑥;  𝛼, 𝜃, 𝜆) = 𝑒−(𝛼𝑥𝜃+𝛼(𝜆𝑥)𝜃)               (7)  and the hazard function is: ℎ(𝑥;  𝛼, 𝜃, 𝜆) = (1 + 𝜆𝜃)𝛼𝜃𝑥𝜃−1                                              (8) Figure 3 illustrates possible shapes of the hazard function of the weighted Weibull distribution for some selected values of the parameters 𝛼, 𝜃  and 𝜆. European Scientific Journal March  2015 edition vol.11, No.9 ISSN: 1857 – 7881 (Print)  e - ISSN 1857- 7431 137  Figure 3: Hazard function of weighted Weibull distribution  Statistical Properties  In this section, the statistical properties of the weighted Weibull distribution is studied. The quantile function, skewness, kurtosis, mode, moment and moment generating function have been derived.  Quantile function and Simulation   Let 𝑄(𝑢), 0 < 𝑢 < 1 denote the quantile function for the weighted Weibull distribution. Then 𝑄(𝑢) is given   by: 𝑄(𝑢) = 𝐹−1(𝑢) = [ln (11 − 𝑢)𝛼(1 + 𝜆𝜃)]1𝜃                                         (9)  In particular, the distribution of the median is: 𝑄(0.5) = [ln 2𝛼(1 + 𝜆𝜃)]1𝜃  To simulate from the weighted Weibull distribution is straight forward. Let 𝑢 be a uniform variate on the unit interval (0,1). Thus by means of the inverse transformation method, we consider the random variable  𝑋 given by: European Scientific Journal March  2015 edition vol.11, No.9 ISSN: 1857 – 7881 (Print)  e - ISSN 1857- 7431 138 𝑋 = [ln (11 − 𝑢)𝛼(1 + 𝜆𝜃)]1𝜃                                                        (10)  Mode     Consider the density of the weighted Weibull distribution given in (5). The mode is obtained by solving 𝑑 ln 𝑓(𝑥)𝑑𝑥= 0 for 𝑥. Therefore the mode at 𝑥 = 𝑥0 is given by: 𝑥0 = [(𝜃 − 1)𝛼(1 + 𝜆𝜃)]1𝜃                                                            (11)  Skewness and Kurtosis  In this study, the quantile based measures of skewness and kurtosis was employed due to non-existence of the classical measures in some cases. The Bowley’s measure of skewness based on quartiles is given by: 𝐵 =𝑄(3/4) − 2𝑄(1/2) + 𝑄(1/4)𝑄(3/4) − 𝑄(1/4) and the Moors’ kurtosis is on octiles and is given by: 𝑀 =𝑄(7/8) − 𝑄(5/8) − 𝑄(3/8) + 𝑄(1/8)𝑄(6/8) − 𝑄(2/8)  where 𝑄(. ) represents the quantile function.  Moment and Moment Generating Function  In this section, the 𝑟𝑡ℎ non central moment and the moment generating function have been derived. Theorem 1. If a random variable 𝑋 has a weighted Weibull distribution, then the 𝑟𝑡ℎ non central moment is given by the following: 𝜇𝑟′ = [1𝛼(1 + 𝜆𝜃)]𝑟𝜃Γ (1 +𝑟𝜃)                            (12)  Proof. 𝜇𝑟′ = ∫ 𝑥𝑟𝑓(𝑥)∞0𝑑𝑥  This implies 𝜇𝑟′ = ∫ 𝑥𝑟(1 + 𝜆𝜃)𝛼𝜃𝑥𝜃−1𝑒−(𝛼𝑥𝜃+𝛼(𝜆𝑥)𝜃)∞0𝑑𝑥           (13) Let 𝑦 = 𝛼𝑥𝜃 + 𝛼(𝜆𝑥)𝜃, 𝑑𝑦 = 𝛼𝜃(1 + 𝜆𝜃)𝑥𝜃−1𝑑𝑥 and 𝑥 = [𝑦𝛼(1+𝜆𝜃)]1𝜃 European Scientific Journal March  2015 edition vol.11, No.9 ISSN: 1857 – 7881 (Print)  e - ISSN 1857- 7431 139 𝜇𝑟′ = ∫ [𝑦𝛼(1 + 𝜆𝜃)]𝑟𝜃𝑒−𝑦𝑑𝑦∞0                                              (14) = [1𝛼(1 + 𝜆𝜃)]𝑟𝜃∫ 𝑦(𝑟𝜃+1)−1𝑒−𝑦𝑑𝑦∞0                               = [1𝛼(1 + 𝜆𝜃)]𝑟𝜃Γ (1 +𝑟𝜃)                                                This completes the proof. If 𝑟 = 1,   𝐸(𝑋) = [1𝛼(1+𝜆𝜃)]1𝜃Γ (1 +1𝜃) If 𝑟 = 2,  𝐸(𝑋2) = [1𝛼(1+𝜆𝜃)]2𝜃Γ (1 +2𝜃)  Therefore the variance is given by 𝑉𝑎𝑟(𝑋) = 𝐸(𝑋2) − {𝐸(𝑋)}2 Theorem 2. Let 𝑋 have a weighted Weibull distribution. The moment generating function  of 𝑋 denoted by 𝑀𝑋(𝑡) is given by: 𝑀𝑋(𝑡) = ∑𝑡𝑖𝑖!∞𝑖=0[1𝛼(1 + 𝜆𝜃)]𝑖𝜃Γ (1 +𝑖𝜃)                              (15)  Proof.  By definition 𝑀𝑋(𝑡) = 𝐸(𝑒𝑡𝑥) = ∫ 𝑒𝑡𝑥∞0𝑓(𝑥)𝑑𝑥  Using Taylor series 𝑀𝑋(𝑡) = ∫ (1 +𝑡𝑥1!+𝑡2𝑥22!+ ⋯ +𝑡𝑛𝑥𝑛𝑛!+ ⋯ ) 𝑓(𝑥)𝑑𝑥∞0 = ∑𝑡𝑖𝐸(𝑋𝑖)𝑖!∞𝑖=0                                                       = ∑𝑡𝑖𝑖!∞𝑖=0[1𝛼(1 + 𝜆𝜃)]𝑖𝜃Γ (1 +𝑖𝜃)                       This completes the proof.  Renyi Entropy  If 𝑋 be a random variable having a weighted Weibull distribution. An important measure of the uncertainty of 𝑋 is the Renyi entropy. The Renyi entropy is defined as: European Scientific Journal March  2015 edition vol.11, No.9 ISSN: 1857 – 7881 (Print)  e - ISSN 1857- 7431 140 𝐼𝑅(𝛿) =11 − 𝛿log[𝐼(𝛿)] where 𝐼(𝛿) = ∫ [𝑓(𝑥)]𝛿𝑑𝑥𝑅, 𝛿 > 0 and 𝛿 ≠ 1. Theorem 3. For a random variable 𝑋 having a weighted Weibull distribution, the Renyi entropy is given by: 𝐼𝑅(𝛿) =11 − 𝛿log [𝜃(1+ 𝜆𝜃)𝛿(𝛼𝜃)𝛿 ∑(−1)𝑖(𝛼𝛿𝜆𝜃)𝑖(𝛼𝛿)−(𝛿𝜃+𝛿+𝑖𝜃+1)𝑖!Γ(𝛿𝜃 − 𝛿∞𝑖=0+ 𝑖𝜃 + 1)] (16)  Where Γ(∙) is the gamma function.  Reliability  Reliability of a component plays a significant role in Stress-Strength analysis of a model. If 𝑋 is the strength and 𝑌 is the stress, the component fails when 𝑋 ≤ 𝑌. Then the estimation of the reliability of the component 𝑅 is Pr (𝑌 < 𝑋). 𝑅 = ∫ 𝑓(𝑥)𝐹(𝑥)𝑑𝑥∞0= 1 − ∫ 𝑓(𝑥)?̅?(𝑥)𝑑𝑥∞0 Theorem 4. If 𝑋 is the strength and 𝑌 is the stress, then the reliability of the component 𝑅 is given by: 𝑅 = 1 − ∑(−1)ℎ𝛼ℎ(1 + 𝜆𝜃)ℎ[𝛼(1 + 𝜆𝜃)]−ℎℎ!∞ℎ=0Γ(ℎ + 1)           (17)  Order Statistics  Let 𝑋(1) denote the smallest of {𝑋1, 𝑋2, … , 𝑋𝑛}, 𝑋(2) denote the second smallest of {𝑋1, 𝑋2, … , 𝑋𝑛}, and similarly 𝑋(𝑘) denote the 𝑘𝑡ℎ smallest of {𝑋1, 𝑋2, … , 𝑋𝑛}. Then the random variables 𝑋(1), 𝑋(2), … , 𝑋(𝑛), called the order statistics of the sample 𝑋1, 𝑋2, … , 𝑋𝑛, has probability density function of the 𝑘𝑡ℎ order statistic, 𝑋(𝑘), as: 𝑔𝑘:𝑛(𝑥) =𝑛!(𝑘 − 1)! (𝑛 − 𝑘)!𝑓(𝑥)[𝐹(𝑥)]𝑘−1[1 − 𝐹(𝑥)]𝑛−𝑘 for 𝑘 = 1, 2, 3, … , 𝑛.  The pdf of the 𝑘𝑡ℎ order statistic is defined as: European Scientific Journal March  2015 edition vol.11, No.9 ISSN: 1857 – 7881 (Print)  e - ISSN 1857- 7431 141 𝑔𝑘:𝑛(𝑥) =𝑛!(𝑘 − 1)! (𝑛 − 𝑘)!(1 + 𝜆𝜃)𝛼𝜃𝑥𝜃−1𝑒−(𝛼𝑥𝜃+𝛼(𝜆𝑥)𝜃)  ×  [1 − 𝑒−(𝛼𝑥𝜃+𝛼(𝜆𝑥)𝜃)]𝑘−1  × [𝑒−(𝛼𝑥𝜃+𝛼(𝜆𝑥)𝜃)]𝑛−𝑘                                            (18)             The pdf of the largest order statistic 𝑋(𝑛) is therefore: 𝑔𝑛:𝑛(𝑥) = 𝑛(1 + 𝜆𝜃)𝛼𝜃𝑥𝜃−1𝑒−(𝛼𝑥𝜃+𝛼(𝜆𝑥)𝜃)×  [1 − 𝑒−(𝛼𝑥𝜃+𝛼(𝜆𝑥)𝜃)]𝑛−1    (19)             and the pdf of the smallest order statistic 𝑋(1) is given by:  𝑔1:𝑛(𝑥) = 𝑛(1 + 𝜆𝜃)𝛼𝜃𝑥𝜃−1𝑒−(𝛼𝑥𝜃+𝛼(𝜆𝑥)𝜃)× [𝑒−(𝛼𝑥𝜃+𝛼(𝜆𝑥)𝜃)]𝑛−1                 (20) Theorem 5. The 𝑟𝑡ℎ non central moment of the 𝑘𝑡ℎ order statistics is given by: 𝜇𝑟′(𝑘:𝑛)=𝑛!(𝑘−1)!(𝑛−𝑘)!∑ ∑ ∑ (𝑘−1𝑗)(−1)𝑗+𝑙+𝑚(𝛼𝑗+𝛼𝑗𝜆𝜃)𝑙[𝛼(𝑛−𝑘)+𝛼𝜆𝜃(𝑛−𝑘)]𝑚𝑙!𝑚!∞𝑗=0∞𝑙=0∞𝑚=0 ×[𝛼(1 + 𝜆𝜃)]−(𝑟+𝑙𝜃+𝑚𝜃𝜃)Γ (𝑟+𝑙𝜃+𝑚𝜃𝜃+ 1)                      (21)   Maximum Likelihood Estimation  In this section, the method of maximum likelihood was considered for the estimation of the parameters of the weighted Weibull distribution. Consider a random sample of size 𝑛, consisting of value 𝑥1, 𝑥2, … , 𝑥𝑛 from the weighted Weibull density 𝑓(𝑥;  𝛼, 𝜃, 𝜆) = (1 + 𝜆𝜃)𝛼𝜃𝑥𝜃−1𝑒−(𝛼𝑥𝜃+𝛼(𝜆𝑥)𝜃)  The likelihood function of the above density is given by: 𝐿(𝒙;  𝛼, 𝜆, 𝜃) = ∏ [(1 + 𝜆𝜃)𝛼𝜃𝑥𝜃−1𝑒−(𝛼𝑥𝜃+𝛼(𝜆𝑥)𝜃)]𝑛𝑖=1 where 𝒙 = [𝑥1, 𝑥2, … , 𝑥𝑛 ]′. The log-likelihood function is given by: ln 𝐿(𝒙;  𝛼, 𝜆, 𝜃) = 𝑛 ln(1 + 𝜆𝜃) + 𝑛 ln 𝛼 + 𝑛 ln 𝜃 + (𝜃 − 1) ∑ ln 𝑥𝑖𝑛𝑖=1                                               −𝛼(1 + 𝜆𝜃) ∑ 𝑥𝑖𝜃𝑛𝑖=1                       (22)    Taking the partial derivatives of the log-likelihood function in (22) with respect to the parameters 𝛼, 𝜆 and 𝜃 yields: European Scientific Journal March  2015 edition vol.11, No.9 ISSN: 1857 – 7881 (Print)  e - ISSN 1857- 7431 142 𝜕 ln 𝐿(𝒙;  𝛼, 𝜆, 𝜃)𝜕𝛼=𝑛𝛼− (1 + 𝜆𝜃) ∑ 𝑥𝑖𝜃𝑛𝑖=1                                       (23) 𝜕 ln 𝐿(𝒙;  𝛼, 𝜆, 𝜃)𝜕𝜆=𝑛𝜃𝜆𝜃−1(1 + 𝜆𝜃)− 𝛼𝜃𝜆𝜃−1 ∑ 𝑥𝑖𝜃𝑛𝑖=1                            (24) 𝜕 ln 𝐿(𝒙;  𝛼, 𝜆, 𝜃)𝜕𝜃=𝑛𝜆𝜃 ln 𝜆(1 + 𝜆𝜃)+𝑛𝜃+ ∑ ln 𝑥𝑖𝑛𝑖=1− 𝛼 ∑[𝑥𝑖𝜃 ln 𝑥𝑖 + (𝜆𝑥𝑖)𝜃 ln(𝜆𝑥𝑖)]𝑛𝑖=1  (25)   Setting equations (23), (24) and (25) to zero and solving them simultaneously yields the maximum likelihood estimates of the three parameters. By taking the second partial derivatives of (23), (24) and (25) the Fisher’s information matrix can be obtained by taking the negative expectations of the second partial derivatives. The inverse of the Fisher’s information matrix is the variance covariance matrix of the maximum likelihood estimators.  Empirical Study  In this section, an empirical study was carried out to investigate the effect of change in the values of the new parameter 𝜆 for 𝛼 = 1.5 and 𝜃 =2.5. Table 1 provides the mean, variance, Bowley’s coefficient of Skewness and Moor’s coefficient of kurtosis. From Table 1, the mean and the variance of the weighted Weibull distribution decreases for an increase in the value of 𝜆. However, the coefficient of skewness and kurtosis are not affected by increase in the value of 𝜆. Table 1: Mean, Variance, Skewness and Kurtosis of weighted Weibull distribution 𝜆 𝛼 = 1.5                        𝜃 = 2.5 Mean Variance Skewness Kurtosis 0.2 0.7490938 0.1027476 0.03727047 0.1043721 0.3 0.7400435 0.1002798 0.03727047 0.1043721 0.4 0.7258902 0.09648081 0.03727047 0.1043721 0.5 0.7068685 0.09149059 0.03727047 0.1043721 1.0 0.5717475 0.059856 0.03727047 0.1043721 1.5 0.4443662 0.03615607 0.03727047 0.1043721 2.0 0.3534343 0.02287265 0.03727047 0.1043721 2.5 0.2903561 0.01543693 0.03727047 0.1043721 3.0 0.2452979 0.03727047 0.03727047 0.1043721 European Scientific Journal March  2015 edition vol.11, No.9 ISSN: 1857 – 7881 (Print)  e - ISSN 1857- 7431 143 Application  In this section, the application of the new weighted Weibull distribution is demonstrated using the lifetime data of 20 electronic components (see Murthy et al., 2004, pp. 83,100). Teimouri and Gupta (2013) studied this data using a three-parameter Weibull distribution. In this study, the weighted Weibull distribution is fitted to this data and the results compared to that of Teimouri and Gupta (2013). The data is shown in Table 2. From Table 3, the Anderson-Darling (AD) statistics revealed that the weighted Weibull fits the data better than the three-parameter Weibull distribution. Table 2: Lifetimes of 20 electronic components 0.03 0.22 0.73 1.25 1.52 1.8 2.38 2.87 3.14 4.72 0.12 0.35 0.79 1.41 1.79 1.94 2.4 2.99 3.17 5.09  Table 3: Estimated Parameters of dataset Distribution Estimated Parameters AD Statistic Weighted Weibull 𝛼 = 0.363        𝜃 = 1.196        𝜆 = 0.233 0.419 Three-parameter Weibull 𝛼 = 1.217    𝛽 = 2.072     𝜇 = −0.008 0.432  Conclusion  A new weighted Weibull distribution based on modified weighted version of Azzalini’s (1985) approach has been proposed. Some important and mathematical properties of the distribution have been derived. An empirical study was carried out to determine the effect of the new parameter on the mean, variance, skewness and kurtosis of the distribution. The application of the new distribution has been demonstrated using real life data. Future works include comparison of the new distribution with other modified weibull distributions, application of the distribution to censored dataset and comparison of different techniques for estimating the parameters of the distribution.  References: Almalki, S. J. and Yuan, J. (2013). The new modified Weibull distribution. Reliability Engineering and System Safety, 111: 164-170. Al-Saleh, J. A. and Agarwal, S. K. (2006). Extended Weibull type distribution and finite mixture of distributions. Statistical Methodology, 3: 224-233. Azzalini, A. (1985). A class of distributions which includes the normal ones. Scandinavian Journal of Statistics, 12: 171-178. Lai, C. D., Xie, M. and Murthy, D. N. P. (2003). A modified Weibull distribution. 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ANOTHER WEIGHTED WEIBULL DISTRIBUTION FROM AZZALINI’S FAMILY

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ANOTHER WEIGHTED WEIBULL DISTRIBUTION FROM AZZALINI’S FAMILY

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