A Comparison of Three Different Procedures for Estimating Variance Components

Transcription

1 International Journal of Statistics and Applications 2016, 6(4): DOI: /j.statistics A Coparison of Three Different Procedures for Estiating Variance Coponents Zakaria A. Abdel Wahed 1, Mohaed S. Abdallah 2,* 1 Faculty of Econoics & Political Science, Cairo University, Egypt 2 Cairo University, Egypt Abstract As a consequence of various theoretical developents, and of iproveents in coputing strategies, restricted axiu likelihood (REML) estiation has becoe a viable procedure for estiating the variance coponents in ixed linear odels. In this article, the procedure of Xu and Atchley (1996) has been extended in such a way that can be applied to any general linear odel. Further, alternative estiator based on epirical Bayes approach has been derived for estiating the rando- effects variance coponents in the light of REML function. Coparison between the proposed estiator and the estiators provided by Xu and Atchley (1996) and Moghtased-Azar et al. (2014) has been coputed under unbalanced nested-factorial odel with two fixed crossed factorial and one nested rando factor. Finally, all the estiators in the vignette are activated by an illustrative exaple. Keywords Epirical Bayes, Negative Estiates, Nested Factorial Design, Restricted Maxiu Likelihood, Siulation Study, Variance Coponents 1. Introduction One of the ost traditional ethods to estiate the variance coponents is REML which is quoted by Thopson (1962). The REML approach ay lead to assive coputations, thus any algoriths were presented over the years to calculate REML estiates. A proper and coprehensive review of these algoriths can be found in Meza (1993) and Matilainen (2014). One of the interesting features of REML is considering the fixed effects as nuisance paraeters in order to take into account the iplicit degrees of freedo associated with the fixed effects. A copeting estiation approach is based on epirical Bayes (EB) principle which, as opposed to REML, considers both the fixed and the rando effects as nuisance paraeters whereas the variance coponents are estiated as hyper paraeters fro the arginal likelihood function which includes only the observed data and the variance coponents. In this article, an attept was ade to produce a new estiator that erges between the REML ethod and the EB approach. It is worthwhile to ention that Coponents of variance estiation can be traced back to the work of astronoer Airy (1861). Since this tie rearkable developents and coparisons have taken place on estiation the variance * Corresponding author: statisticss.2010@gail.co (Mohaed S. Abdallah) Published online at Copyright 2016 Scientific & Acadeic Publishing. All Rights Reserved coponents either fro theoretical perspective, epirical view or both. Consequently a plethora of variance coponents estiation were proposed resulting in extensive publications of books and working papers in this area. A proper survey review for these ethods can be found in Sahai (1979), Khuri and Sahai (1985), Khuri (2000) and Sahia and Khurshid (2005). Due to the coplexity of the variance coponents odels, it ay be intractable to conduct any theoretical coparison aong variance coponents estiators even for the siplest cases of the linear ixed odel with unbalanced data. Consequently a nuber of attepts were perfored to study the statistical properties of the variance coponents estiators using Monte Carlo siulation under different types of odels. For instance, based on crossed odels: Swallow and Monahan (1984) ade a coparison aong ANOVA, MLE, REML, and MINQUE ethods, Khattree (1998) copared between his proposed non-negative variance coponents estiator based on siple odification of Henderson's ANOVA ethod and the original Henderson's ANOVA ethod, Lee and Khuri (2000) used the Epirical Quantile Dispersion Graph (EQDG) to ake a coparison between the ANOVA and ML estiation ethods, Subraani (2012) introduced a new procedure to estiate the variance coponents in the light of MINQUE approach and copared with ANOVA and MINQUE ethod, and Chen and Wei (2013) derived paraetric epirical Bayes estiators and investigated the superiorities of their estiator over ANOVA ethod. On the other hand, coparisons based on nested odels: Sahai (1976)

2 242 Zakaria A. Abdel Wahed et al.: A Coparison of Three Different Procedures for Estiating Variance Coponents copared aong ANOVA, MLE and REML ethods, Rao and Hecker (1997) provided a new estiator based on weighted analysis of eans (WAM) estiator which utilizes prior inforation for estiating the variance coponents, further they copared nuerically between their estiator and other traditional variance coponents ethods, Jung et al. (2008) copared between ANOVA and MLE ethods based upon the EQDG. Finally El Leithy et al. (2016) suggested a new estiator based on the estiator of Subraani (2012) and ade a coparative study between the proposed estiator and other coon estiators using nested factorial odels with two crossed factors and one nested factor. This article is structured as follows: Section 2 will explain in details the variance coponent estiator proposed by Moghtased-Azar et al. (2014) which is donated as odified REML (MREML) estiator. In Section 3, the variance coponent estiator based on EB aspect proposed by Xu and Atchley (1996) will be discussed. A proposed variance coponent estiator is introduced in Section 4 which is called odified epirical Bayes (MEB) estiator. Section 5 includes the Monte Carlo results using nested-factorial odel. Section 6 includes an illustrative application for all the preceding estiators. Finally, soe conclusions about this work are given in the last section. 2. MREML Method Consider the variance coponents odel Y = Xβ + Z 1 δ 1 + Z 2 δ 2 + Z r 1 δ r 1 + Z r δ r (2.1) where Y is an n 1 vector of observations, X is an n atrix with known constants, β is an 1 vector of fixed (unknown) paraeters, Z i is an n c i atrix of known constants and δ i is c i 1 rando vector which has ultivariate noral distribution with zero ean and covariance atrix σ 2 ii ci i = 1 r. Further it is assued that δ i and δ j i j are uncorrelated. Model (2.1) can be expressed in a copact for as: Y = Xβ + Zδ (2.2) where Z = [Z 1 Z 2 Z r ] and δ = [δ 1δ 2 δ r]. The odel (2.2) is called a ixed linear, odel. If r = 1, it becoes a fixed odel and if = 1 it becoes a rando odel. Thus generally we have E Y = Xβ and D Y = r i=1 σ 2 iv i, where V i = Z i Z i, D is called the dispersion atrix and the paraeters σ 2 1, σ 2 2,, σ 2 r are the unknown variance coponents whose values should be estiated. Since the norality distribution is assued, thus it is acceptable to operate distribution-based ethods. The preferred paraetric ethod for estiating variance coponents is REML. The original reference to REML is the article by Thopson (1962). Theoretically, REML can be illustrated as assuing K n x,n be a full rank atrix, where x is the rank of X, such that KX = 0, 1 then the likelihood of KY can be forulated as: L σ\y KDK 1 2 exp( 1 2 (Y K (KDK ) 1 KY)) (2.3) where σ = [σ 2 1σ 2 2 σ 2 r] is a vector of the unknown variance coponents. The log likelihood of KY becoes: ln L (σ\y) 1 2 ln KDK 1 2 (Y K (KDK ) 1 KY) in order to obtain REML estiates, it is required to take the partial derivatives of ln(l(σ\y)) with respect to σ then setting to zero, we obtain ln(l(σ\y)) σ 2 = 1 i 2 (tr K KDK 1 KV i + Y K (KDK ) 1 KV i K (KDK ) 1 KY) = 0 using the lea given in Searle et al.(1992) that states: hence we will get K (KDK ) 1 K = P, where P = D 1 D 1 X(X D 1 X) 1 X D 1 ln(l(σ\y)) σ 2 i = tr PV i Y PV i PY = 0 i = 1 r (2.4) It is obvious that we have r equations in r unknowns σ. In soe cases these equations can be siplified to yield a closed for. Yet, in alost cases nuerical algoriths have to be used to solve the equations, it should also be noted that the syste of equations in (2.4) does not involve the eleents of K, which eans no atter what their values are, the sae result will be reached (see Searle et al. (1992)). Nuerically, Rao and Kleffe (1988) affired that REML solutions are equivalent to other variance coponents estiators (e.g. iterated MINQE, iterated BIQUE and ANOVA) under suitable conditions, weakly consistent, asyptotically efficient and asyptotically noral with zero ean and variance atrix equals to the inverse of the restricted inforation atrix. Cressie and Lahiri (1993) studied the consistency and asyptotic norality (CAN) of REML estiates in the light of the results reached by sweeting (1980), and applied REML to census undercount in the USA. Jiang (1996) studied the REML estiators in the absence of norality, boundedness of and prior specification of the structure of the odel. He resorted to deal with REML estiates as a kind of M-estiates in order to reach to the sae REML equations. Further, he proved rigorously that the REML estiates are consistent under odels that are asyptotically identifiable infinitely and whether the nondegenerate condition is added, then REML estiates are asyptotically noral. Finally, he showed that in the light of the nondegenerate, the REML and ANOVA estiates are consistent and asyptotically noral for all unconfounded

3 International Journal of Statistics and Applications 2016, 6(4): balanced ixed odels. The ain drawback concerning the REML technique is that the solution in (2.4) can be negative, which it is not allowed in the real life. This dilea has been resolved by Moghtased-Azar et al. (2014). Moghtased-Azar et al. (2014) claied that using any positive valued functions can guarantee the non-negativity of the variance coponents estiates, thus they decided to depend on the exponential function as its range is positive everywhere. Hence obtaining MREML required first to re-paraetrize ln(l(σ\y)) as replacing σ 2 i with exp(α i ), then axiizing ln(l(σ\y)) with respect to α i instead of σ 2 i. After little algebra (2.4) can be reforulated in ters of α i as: g α = ln(l(α\y)) α i = tr P V i Y P V i P Y = 0 i = 1 r (2.5) where α = α 1 α 2 α r is a vector of the unknown paraeters, P = D 1 D 1 X X D 1 X 1 X D 1 = K (KD K ) 1 K and D r = i=1 exp(α i )V i. As explained above, we have r equations in r unknowns α i, solving these nonlinear equations through any iterative algorith, then the estiated variance coponents can be calculated by σ = exp(α). Fortunately Moghtased-Azar et al. (2014) derived an efficient algorith in order to estiate α i via linearizing (2.5) using Taylor expansion with first order as follows: g α = g α g α + α α=α α α = 0 (2.6) where α is the prior value of the α. The only unknown ter in (2.6) is the first derivative of g α which can be obtained as: After tedious algebra g i α g i α g i α can be expressed as: = (tr P V i Y P V i P Y) = (tr K (KD K ) 1 KV i (Y K (KD K ) 1 KV i K (KD K ) 1 KY) j = 1 r (2.7) = 2exp(α j )( Y P V j P V i P Y) exp(α i )tr P V j P V i i, j = 1 r g α For siplification, Moghtased-Azar et al. (2014) preferred to deal with E( ), which can described as: E g i α It can easily be proved that: = 2exp(α j )E( Y P V j P V i P Y) exp(α i )tr P V j P V i E Y P V j P V i P Y = tr K KD K 1 KV j P V i K KD K 1 KD = tr P V j P V i hence E g i α will becoe: E g i α = 2 exp α j tr P V j P V i exp α i tr P V j P V i = exp α i tr P V j P V i (2.8) Recall Taylor equation which can be rewritten as: where H i,j = E g i α g α + H α α = 0 Hα = Hα g α α = α H 1 g α (2.9). Consequently getting on MREML based on the following steps: 1) Select initial values for α i, in order to calculate g α using (2.5). 2) Calculate the H atrix using (2.8). 3) Evaluate α using (2.9). Then copute σ = exp(α). 4) Update the initial values in the step (1) with α evaluated fro step (3) and repeat the circle until the convergence achieved. Actually Moghtased-Azar et al. (2014) proved atheatically that MREML is an unbiased estiator and has the sae variance as REML, further they showed using an illustrated exaple that the perforance of above algorith is satisfying and converges quickly.

4 244 Zakaria A. Abdel Wahed et al.: A Coparison of Three Different Procedures for Estiating Variance Coponents 3. EB Method In the Bayesian approach, all the paraeters are regarded as rando in the sense that all uncertainty about the should be expressed in ters of a probability distribution function. The basic paradig of Bayesian statistics involves a choice of a joint prior distribution of all paraeters of interest that could be based on objective evidence or subjective judgent or a cobination of both. Generally speaking, the Bayesian approach can be divided into two principal parts. The first one is known as hierarchical Bayesian odels which the linear ixed odel is considered as a ulti-stage hierarchy and all the paraeters of the odel including variance coponents should have prior distributions, for ore details see Jiang (2007). The second case is known as epirical Bayes or arginal likelihood function that all the nuisance paraeters, fixed and rando effects, should be integrated out which yields a likelihood function that includes only the observed data and the variance coponents, thus one can estiate the variance coponents, as in the classical situations, through axiizing the arginal likelihood with respect to the unknown the variance coponents (for ore details see Searle et al. (1992)). Harville (1977) proved that REML can be considered as an epirical Bayes estiator. He showed that the REML can be derived as the arginal likelihood when the fixed effects are integrated out under a non-inforative or flat prior distribution. He used the unifor distribution (-,), f β = 1, as a prior distribution for the fixed effects β, however the rando effects δ i is defined as in (2.1). The arginal likelihood y σ can be expressed as: where: y σ = L y β, δ, σ exp( 1 2 (Y Xβ Zδ) (Y Xβ Zδ)), π δ σ R 1 2exp( 1 2 ( δ R 1 δ)), and R = L y β, δ, σ π δ σ dβ σ 2 1I c σ 2 2I c σ 2 ri cr Since the variance coponents obtained in (2.4) is equivalent to the variance coponents obtained using y σ, hence the latter suffers also fro the negativity proble. Xu and Atchley (1996) proposed a new idea that can be adopted for the estiating non-negative variance coponents based on epirical Bayes using Monte-Carlo integration algorith. In this context, the idea will be generalized through the following steps: 1- Express the rando effect vector δ i as a linear transforation of standard noral deviates which can be expressed as: dδ. δ i = S i σ i, where S i ~N ci 1(0,1) (3.1) 2- Use L( σ Y) in (2.3) to derive a new arginal likelihood function in the light of (3.1) as given below: 1 y σ exp( 1 2 ((Y ZSσ ) P (Y ZSσ )))exp( 1 2 (S S))dS where S = S 1, S 2, S r, σ = [σ 1 σ 2 σ r ] and P = I X(X X) X. 3- Maxiize y σ with respect to σ, then σ = σ 2. Which iplies that σ is positive under all conditions. Since S is copletely known following standard noral distribution, one can easily generate a huge set of standard noral distribution. Therefore Xu and Atchley (1996) suggested replacing the integrations in y σ with the suations to siplify the analysis which yields to: y σ exp 1 2 Y Z 1S 1z σ 1 Z 2 S 2z σ 2 Z r S rz σ r P Y Z 1 S 1z σ 1 Z 2 S 2z σ 2 Z r S rz σ r One should notice that all S iz can be obtained fro generating standard noral saples, they stated that as, y σ y σ, so they recoend to set = The axiu likelihood solution of σ required to take the partial derivatives of the ln( y σ ) with respect to σ as follows: 1 Xu and Atchley (1996) discussed ultiple arginal likelihood functions, yet we confine ourselves to select which based only on REML.

5 International Journal of Statistics and Applications 2016, 6(4): ln( y σ ) σ = where p z S 1z Z 1 P Y p z S 2z Z 2 P Y 1 y σ ( y σ ) σ = p z S 1z Z 1 P Z 1 S 1z σ 1 p z S 2z Z 2 P Z 1 S 1z σ 1 p z S 1z Z 1 P Z r S rz σ r p z S 2z Z 2 P Z 2 S 2z σ 2 p z S 2z Z 2 P Z r S rz σ r p z S rz Z r P Y p z S rz Z r P Z 1 S 1z σ 1 p z S rz Z r P Z r S rz σ r = 0 exp 1 2 Y Z 1S 1z σ 1 Z 2 S 2z σ 2 Z r S rz σ r P Y Z 1 S 1z σ 1 Z 2 S 2z σ 2 Z r S rz σ r p z = y σ The ML of σ ay be written as a linear syste by: Where A = p z S 1z Z 1 P Z 1 S 1z p z S 2z Z 2 P Z 1 S 1z Aσ = C p z S 1z Z 1 P Z r S rz p z S 2z Z 2 P Z 2 S 2z p z S 2z Z 2 P Z r S rz p z S rz Z r P Z 1 S 1z Consequently σ can be calculated as: p z S rz Z r P Z r S rz σ = A 1 C and C = p z S 1z Z 1 P Y p z S 2z Z 2 P Y p z S rz Z r P Y Although algebraically Xu and Atchley (1996) estiator sees to be tedious, it is conceptually very siple, easy to progra and avoids inverting large atrix as opposed to MREML estiator which requires to copute the inverse of D n n during calculating P. 4. MEB Method One can derive a new estiator based on EB estiator. Recall that y σ in ters of δ that can be expressed as: y σ R 1 2exp( 1 2 ((Y Zδ) P (Y Zδ)))exp( 1 2 (δ R 1 δ))dδ Our idea is to solve the above integration analytically, then estiate σ using the algorith explained in Moghtased-Azar et al. (2014). Fortunately the above integration has explicit for which enables us to obtain MEB as follows: y σ R 1 2 exp 1 2 Y Zδ P Y Zδ + δ R 1 δ dδ R 1 2exp( 1 2 (Y P Y 2δ Z P Y + δ Z P Zδ + δ R 1 δ))dδ R 1 2exp( 1 2 (Y P Y 2δ Z P Y + δ (Z P Z + R 1 )δ))dδ Let us define A = (Z P Z + R 1 ), hence we can get:

6 246 Zakaria A. Abdel Wahed et al.: A Coparison of Three Different Procedures for Estiating Variance Coponents y σ R 1 2exp( 1 2 Y P Y + δ A 1 Z P Y A δ A 1 Z P Y Y P ZA 1 Z P Y )dδ R 1 2exp( 1 2 Y P Y Y P ZA 1 Z P Y + δ A 1 Z P Y A δ A 1 Z P Y )dδ R 1 2 exp 1 2 Y P Y Y P ZA 1 Z P Y exp( 1 2 δ A 1 Z P Y A δ A 1 Z P Y ) dδ R Z P Z + I 1 2exp( 1 2 Y P Y Y P ZA 1 Z P Y ) hence ln( y σ ) can be presented as: ln y σ 1 2 (ln R Z P Z + I Y P P ZA 1 Z P Y) Since R is the variance coponents of δ, thus it can be rewritten as: where R 1 = I c R = σ 2 1R 1 + σ 2 2R 2 + σ 2 rr r, R 2 = I c R r = I cr For preventing the negative values of σ, we resort to the idea of Moghtased-Azar et al. (2014) to treat this dilea as replacing σ with exp(α). The axiu likelihood solution of α required to take the partial derivatives of the ln y α with respect to α as follows: ln y α α i = exp(α i )(tr R Z P Z + I 1 R i Z P Z + Y P ZA 1 R 1 R i R 1 A 1 Z P Y ) = 0 For siplicity we define: g α = tr R Z P Z + I 1 R i Z P Z Y P ZA 1 R 1 R i R 1 A 1 Z P Y (4.1) According to (2.6), Taylor expansion with first order ay be given as: where g α = g α + g α α g i α The first ter in (4.2) can be expressed as: tr R Z P Z + I 1 R i Z P Z The second ter in (4.2) can be derived as: Y P ZA 1 R 1 R i R 1 A 1 Z P Y α=α = tr R Z P 1 Z+I Ri Z P Z α α Y P ZA 1 R 1 R i R 1 A 1 Z P Y (4.2) = exp(α j )tr R Z P Z + I 1 R j Z P Z R Z P Z + I 1 R i Z P Z = Y P Z RZ P Z + I 1 R i Z P Z R + I 1 Z P Y = exp(α j )[Y P Z RZ P Z + I 1 R j Z P Z RZ P Z + I 1 R i Z P Z R + I 1 Z P Y + Y P Z RZ P Z + I 1 R i Z P Z R + I 1 Z P Z R j Z P Z R + I 1 Z P Y] Since the outcoe is scalar, hence: Y P ZA 1 R 1 R i R 1 A 1 Z P Y = 2exp(α j )(Y P Z RZ P Z + I 1 R i Z P Z R + I 1 Z P Z R j Z P Z R + I 1 Z P Y) hence g i α will becoe:

7 International Journal of Statistics and Applications 2016, 6(4): g i α = 2exp(α ) Y P Z RZ P Z + I 1 R i Z P Z R + I 1 Z P Z R j Z P Z R + I 1 Z P Y j exp(α j )tr R Z P Z + I 1 R j Z P Z R Z P Z + I 1 R i Z P Z (4.3) Recall Taylor equation which can be rewritten as: g α + H α α = 0 H α = Hα g α α = α H 1 g α (4.4) where H i,j = g i α. Consequently steps to get MEB are suarized as following: 1) Select initial values for α i, in order to calculate g α using (4.1). 2) Calculate the H atrix using (4.3). 3) Evaluate α using (4.4). Then copute σ = exp(α). 4) Update the initial values in the step (1) with α evaluated fro step (3) and repeat the circle until the convergence achieved. 5. Siulation Study Since it ay be intractable to do any theoretical coparisons about the perforance of the preceding estiators, thus one has to resort to copare through Monte Carlo siulation. Monte Carlo siulation is now a uch-used scientific tool for probles that are analytically intractable and for which experientation is too tie-consuing, costly, or ipractical. It depends basically on generating artificial rando sapling any ties, 1000 ties for instance, in order to estiate the statistical odels and the atheatical functions. Even though, siulation also has disadvantages; it can require huge coputing resources, it doesn't give exact solutions, and results are only as good as the odel and inputs used. Typically coparisons between the variance coponents estiators should be conducted under hypothetical odel. Following to Melo et al. (2013) and El Leithy et al. (2016), nested factorial design with two crossed factors and one nested factor is adopted in this context in order to identify the behavior of variance coponents estiators which can be described as: y abcd = α a + β b + γ c(a) + αβ ab + βγ bc(a) + ε abcd a = 1.. I; b = 1.. J; c = 1.. K a ; d = 1.. n abc, where α a is the effect of the a level of factor A, β b is the effect of the b level of factor B, γ c(a) is the effect of the c level of factor C nested within the a level of A, αβ ab is the interaction effect between the factor A and B, βγ bc(a) is the interaction effect between the factor B and C and ε abcd is a rando experiental error. It is assued that all the effects in the odel are fixed paraeters except γ c(a), βγ bc(a) and ε abcd are norally independently distributed such that: γ c(a) ~N(0, σ 2 1), βγ bc(a) ~N 0, σ 2 2 and ε abcd ~N(0, σ 2 3). The ain reason for selecting the above odel is due to widely usage in the real life, as nested factorial odels have the advantage that is suitable for the experients are either with crossing or nesting factors or both. Since the fixed effects are out of our interest, thus one can fix all the fixed paraeters at one. Conversely, the coparison process requires to be conducted under a variety of variance coponents configurations, difference of ibalance degrees and ultiple saple sizes. Following to El Leithy et al. (2016), Table (1). displays the variance coponents values used in the siulation. The easure which is introduced by Ahrens and Pincus (1981) can be reliable for reflecting the ibalance effect of n abc which can be forulated as: 1 φ = n abc a b c n where n is the grand saple size and is the grand nuber of the cells obtained by J a K a, where K a is the levels nuber of the nested factor within the factor a. Table (2). presents the patterns of ibalance according to different saple sizes throughout the siulation. For each variance coponents configuration and pattern of ibalance cobination, 1000 independent rando saples were generated, the estiated bias, MSE and the average of the required iterations for the convergence are shown in Table (3). Since such these estiators are sensitive to the starting points, alost unbiased estiator (AUE) which proposed by Horn et al. (1975) was considered as an initial value in our analysis. The reasons for depending on AUE: 1- Its calculations are siple as it can be expressed in a closed for, 2- It always gives non-negative values. For ore details about AUE principle see Searle et al. (1992). According to Table (3), a nuber of conclusions is drawn fro the results for all the patterns and designs which are suarized in the following points: a) It is observed that the three types of the estiators are very close with respect to the nuber of the iterations particularly at increasing saple size. This result can be explained in light of considering AUE as a priary point which enables the convergence to be achieved rapidly. Generally speaking, EB has the least nuber of iterations across all the cases. b) It is clear the superiority of MEB over all the estiators in ters of MSE criterion that across alost cases MEB has MSE less than either MREML, EB or both. Whereas MREML estiator has least bias aong the preceding estiators. 2

8 248 Zakaria A. Abdel Wahed et al.: A Coparison of Three Different Procedures for Estiating Variance Coponents c) The saple size and the ibalance rate have substantial effect on the behavior of all the estiators, as either increasing the sall size or reducing the ibalance rate yield a significant iproveent in the two easures of the perforance. Furtherore, there is an interaction effect between the saple size and the ibalance rate as the effect of the ibalance rate is downward at high level of the saple size. 6. Illustrative Exaple In this part we will just exaine the applicability of MREML, EB, and MEB estiators through conducting these estiators on practical case especially large dataset. The application within Inforation and counication technologies (ICTs) field is provided and conducted via nested factorial odel. ICT are used frequently by Private enterprises to iprove their productivity and copetitiveness in the arketplace. Various kinds of ICT infrastructure such as coputer usage, internet usage and obile usage help firs in all sectors to anage their resources ore efficiently. Consequently, it is useful to consider the factors affecting the expenditures of ICT by Private sectors. Accordingly, The Ministry of Counication and Inforation Technology (MCIT) conducted a survey in 2012 to easure the ICT usage in the private sectors, a proportional saple with size 5,000 enterprises was selected to reflect ICT usage of all Egyptian enterprises. In this context, we will focus only on the ICT expenditures, the location, the region and the International Standard Industrial Classification (ISIC) V.3. of the firs. It is assued that the ICT expenditures can be ostly affected by the location, the region and the ISIC of the firs. We used the nested factorial design to exaine this effect. The raw data are suarized by Table (4) and Figure (1) (see Annex). For the sake of copleteness, we write the odel equation with the definition of the fixed and rando factors in ters of the preceding application as follows: y abcd = α a + β b + γ c(a) + αβ ab + βγ bc(a) + ε abcd a = 1.. 4; b = 1.. 6; c = 1.. [1; 2; 2; 2]; d = 1.. n abc, where y abcd is the dependent variable which is the ICT expenditure for the private sectors after operating the Logarith transforation two ties to guarantee the norality assuption, α a is the effect of the Region factor, β b is the effect of the ISIC factor, γ c(a) is the effect of the Location factor nested within the Region factor, αβ ab is the interaction effect between the Region factor and ISIC factor, βγ bc(a) is the interaction effect between the ISIC factor and the Location factor and ε abcd is a rando experiental error. Since our saple did not include all ISIC activities, it is assued that all the effects in the odel are fixed paraeters except β b, αβ ab, βγ bc(a) and ε abcd are norally independently distributed such that: β b ~N 0, σ 2 b, αβ ab ~N 0, σ 2 ab, βγ bc a ~N 0, σ 2 bc a and ε abcd ~N(0, σ 2 ε). The REML, EB and MEB estiates are calculated for σ 2 b, σ 2 ab, σ 2 bc(a) and σ 2 ε and given in Table (5). Since testing hypotheses is beyond the scope of this study, it ay interest to take up this topic in the near future. In the light of Table (5), the variance coponents estiates corresponding to all rando factors are estiated by MREML, EB and MEB respectively. As excepted MEB estiator required large nuber of iterations to converge. In our view, the reason for this point that it is required to siplify equation (4.3) as Moghtased-Azar et al. (2014) done. One can also observe that the σ 2 ε is relatively large which eans that the odel needs other factors to explain the variability of the ICT expenditure for the private sectors i.e. the capital size, the nuber of the eployees..etc. It is easily to reark that since the high degree of unbalancedness of our data, the variance coponents look soehow different across the studied ethods nonetheless the saple size is large. 7. Conclusions In this article, a new estiator based on epirical Bayes principle is introduced for estiating the variance coponents in the ixed linear odel. The ai of this article is to evaluate the perforance of the proposed estiator relative to other two (MREML and EB) estiators via siulation studies. The odel we used is the nested-factorial odel with two fixed crossed factorial and one nested rando factor under regularity assuptions. Three criteria bias, MSE and nuber of iterations are used to show the perforance of all the estiators under the study. Fro the nuerical analysis, we have found that the proposed estiator has desirable properties in ters of MSE, however MREML estiator ay be appropriate estiators with respect to bias view, and further EB estiator can be considered as a suitable estiator for handling large data sets since it required low nuber of iterations to converge. ACKNOWLEDGMENTS The authors would like to express their heartiest thanks to the editor and the two referees for careful reading and for their coents which greatly iproved the article.

9 International Journal of Statistics and Applications 2016, 6(4): Annex Table (1). Variance Coponents Configurations Used in the Siulation σ 2 1 σ 2 2 σ 2 3 V V V V V V Table (2). The Patterns of Ibalance Rate for Each Saple Size Used in the Siulation n i j K i n ijk φ P ,2 3,,3 100% P ,2 2,5;3,2;5,3;3,1 83% P ,2 1,2;1,2;2,8;7,1 56% P ,2,3 3,,3 100% P ,2,3 2,2;2,5;2,5;4,4;3,3;2,2 87% P ,2,3 1,1;2,5;2,1;1,10;3,1;7,2 53% P ,3,2 3,,3 100% P ,3,2 2,2,4;4,2,4;2,2,4;4,3,6;3,3,4;3,3,2;3,2,1 87% P ,3,2 2,1,1;3,2,10;8,1,2;1,1,3;3,3,2;9,3,3;3,1,1 57% Figure (1). Histogra of the Logarith of the Annual Expense with the Norality Test

10 250 Zakaria A. Abdel Wahed et al.: A Coparison of Three Different Procedures for Estiating Variance Coponents Table (3). Coparison of MREML, EB and MEB Estiators Based on Copound Absolute Bias, Copound MSE and Average of the Iterations MREML EB MEB absolute bias MSE Average of the iterations absolute bias MSE Average of the iterations absolute bias MSE Average of the iterations P1 P2 P3 P4 P5 P6 P7 V V V V V V V V V V V V V V V V V V V V V V V V V V V V V V V V V V V V V V V V V V

11 International Journal of Statistics and Applications 2016, 6(4): MREML EB MEB absolute bias MSE Average of the iterations absolute bias MSE Average of the iterations absolute bias MSE Average of the iterations P8 P9 V V V V V V V V V V V V Table (4). Average of Annual Expense on ICT for Egyptian Private Enterprises Region URBAN LOWER UPPER BORDER ISIC / Location urban urban rural urban rural urban Rural Manufacturing Construction Whole Sale & Retail Trade Hotels & Restaurants Transport, Storage, Counication Financial Interediation Table (5). Estiates of the Rando Factors MREML EB MEB Estiates Iterations Nuber Estiates Iterations Nuber Estiates Iterations Nuber σ 2 b σ 2 ab σ 2 bc(a) σ 2 ε Analysis REFERENCES [1] Ahrens, H. and Pincus, R. (1981), On Two Measures of Unbalanceness in a One-Way Model and Their Relation to Efficiency. Bioetrics [2] Airy, G. B. (1861). On the Algebraical and Nuerical Theory of Errors of Observations and the Cobinations of Observations. MacMillan, London. [3] Cressie. N. and Lahiri. S. (1993). The Asyptotic Distribution of REML Estiators. Journal of Multivariate [4] Chen. L. and Wei. L. (2013). The Superiorities of Epirical Bayes Estiation of Variance Coponents in Rando Effects Model. Counications in Statistics Theory and Methods.42: [5] Diffey. S. Welsh. A. Sith. A. Cullis. B. (2013). A faster and coputationally ore efficient REML (PX) EM algorith for linear ixed odels. Centre for Statistical & Survey Methodology. Working Paper Series. University of Wollongong. [6] El Leithy, H. Abdel-Wahed, Z. and Abdullah, M. (2016). On non-negative estiation of variance coponents in ixed

12 252 Zakaria A. Abdel Wahed et al.: A Coparison of Three Different Procedures for Estiating Variance Coponents linear odels. Journal of Advanced Research [7] Harville. D. (1977). Maxiu-likelihood approaches to variance coponent estiation and to related probles. JASA [8] Horn. S. Horn. R. and Dunca. D. (1975). Estiation Heteroscedastic Variances in Linear Model. JASA.70: [9] Jiang. J. (1996). REML Estiation: Asyptotic Behavior And Related Topics. The Annals of Statistics [10] Jiang. J. (2007). Linear and Generalized Linear Mixed Models and Their Applications. Springer Science +Business Media. [11] Jung. C. Khuri. I. Lee. J. (2008). Coparison of Designs for the Three Folds Nested Rando Model. Journal of Applied Statistics [12] Khuri. R. (1998). Soe practical estiation procedures for variance coponents. Coputational Statistics & Data Analysis [13] Khuri. A. and Sinha. B. (1998). Statistical Tests for Mixed Linear Models. John Wiley & Sons. New York. [14] Khuri. A. and Sahai. H. (1985). Variance Coponents Analysis: A Selective Literature Survey. International Statistical Review [15] Khuri. A. (2000). Designs for Variance Coponents Estiation: Past and Present. International Statistical Review.68: [16] Lee. J. and Khuri. A. (2000). Quantile Dispersion Graphs for the Coparison of Designs for a Rando Two-Way Model. Journal of Statistical Planning and Inference [17] Matilainen. K. (2014). Variance coponent estiation exploiting Monte Carlo ethods and linearization with coplex odels and large data in anial breeding. Ph.D. thesis. Faculty of Agriculture and Forestry. University of Helsinki. Finland. [18] Melo. S. Garzón. B. Melo. O. (2013). Cell Means Model for Balanced Factorial Designs with Nested Mixed Factors. Counications in Statistics Theory and Methods [19] Meza. M. (1993). Estiation of Variance Coponents And Diagnostic Analysis In Unbalanced Mixed Linear Models. Ph.D. Thesis. Texas University. USA. [20] Moghtased-Azar. K. Tehranchi. R. and Airi-Sikooei. A. (2014). An Alternative Method for Non-Negative Estiation of Variance Coponents. Journal Geodesy [21] Qie. W. and Xu. C. (2009). Evaluation of a New Variance Coponents Estiation Method Modified Henderson s Method 3 With the Application of Two Way Mixed Model. Departent of Econoics and Society, Dalarna University College. [22] Rao. C. and Kleffe. J. (1988). Estiation of Variance Coponents and Applications. North-Holland, Asterda. New York. [23] Rao. S. Heckler. E. (1997). The Three-Fold Nested Rando Effects Model. Journal of Statistical Planning and Inference [24] Sahai, A. (1976). A Coparison of Estiators of Variance Coponents in the Balanced Three-Stage Nested Rando Effects Model Using Mean Squared Error Criterion. JASA [25] Sahai, A. (1979). Bibliography On Variance Coponents. International Statistical Review [26] Sweeting. T. (1980). Unifor Asyptotic Norality of The Maxiu Likelihood Estiator. Ann. Statistics [27] Sahai. H. Khurshid. A. (2005). A Bibliography On Variance Coponents An Introduction And An Update: Statistica Applicata 17, [28] Searle. S. Casella. G. and McCulloch. C. (1992). Variance Coponents; John Wiley & Sons. New York. [29] Subraani. J. (2012). On Modified Miniu Variance Quadratic Unbiased Estiation (MIVQUE) of Variance Coponents in Mixed Linear Models. Model Assisted Statistics and Applications [30] Swallow. H. and Monahan. F. (1984). Monte-Carlo Coparison of ANOVA, MINQUE, REML and ML Estiators of Variance Coponents. Technoetrics [31] Thopson. A. (1962). The Proble of Negative Estiates of Variance Coponents. The Annals of Matheatical Statistics [32] Wei. L. Ding. X. (2003). On epirical Bayes estiation of variance coponents in rando effects odel. Journal of Statistical Planning and Inference [33] Xu. S. Atchley. WR. (1996). A Monte-Carlo algorith for axiu likelihood estiation of variance coponents. Genetics Selection Evolution, BioMed Central