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HindawiComplexityVolume 2020, Article ID 8941652, 13 pageshttps://doi.org/10.1155/2020/8941652Research ArticleA Penalized h-Likelihood Variable Selection Algorithm forGeneralized Linear Regression Models with Random EffectsYanxi Xie , Yuewen Li , Zhijie Xia, Ruixia Yan, and Dongqing LuanSchool of Management, Shanghai University of Engineering Science, Shanghai 201620, ChinaCorrespondence should be addressed to Yuewen Li; sues0305@126.comReceived 31 July 2020; Revised 28 August 2020; Accepted 6 September 2020; Published 15 September 2020Academic Editor: Shuping HeCopyright 2020 Yanxi Xie et al. This is an open access article distributed under the Creative Commons Attribution License,which permits unrestricted use, distribution, and reproduction in any medium, provided the original work is properly cited.Reinforcement learning is one of the paradigms and methodologies of machine learning developed in the computational intelligence community. Reinforcement learning algorithms present a major challenge in complex dynamics recently. In theperspective of variable selection, we often come across situations where too many variables are included in the full model at theinitial stage of modeling. Due to a high-dimensional and intractable integral of longitudinal data, likelihood inference iscomputationally challenging. It can be computationally difficult such as very slow convergence or even nonconvergence, for thecomputationally intensive methods. Recently, hierarchical likelihood (h-likelihood) plays an important role in inferences formodels having unobservable or unobserved random variables. This paper focuses linear models with random effects in the meanstructure and proposes a penalized h-likelihood algorithm which incorporates variable selection procedures in the setting of meanmodeling via h-likelihood. The penalized h-likelihood method avoids the messy integration for the random effects and iscomputationally efficient. Furthermore, it demonstrates good performance in relevant-variable selection. Throughout theoreticalanalysis and simulations, it is confirmed that the penalized h-likelihood algorithm produces good fixed effect estimation resultsand can identify zero regression coefficients in modeling the mean structure.1. IntroductionReinforcement learning is specified as trial and error (variation and selection and search) plus learning (associationand memory) in Sutton and Barto [1]. Traditional variableselection procedures, such as LASSO in Tibshirani [2] andOMP in Cai and Wang [3], only consider the fixed effectestimates in the linear models in the past literature. However, in real life, a lot of existing data have both the fixedeffects and random effects involved. For example, in theclinic trials, several observations are taken for a period oftime for one particular patient. After collecting the dataneeded for all the patients, it is natural to consider randomeffects for each individual patient in the model setting since acommon error term for all the observations is not sufficientto capture the individual randomness. Moreover, randomeffects, which are not directly observable, are of interest inthemselves if inference is focused on each individual’s response. Therefore, to solve the problem of the random effectsand to get good estimates, Lee and Nelder [4] proposedhierarchical generalized linear models (HGLMs). HGLMsare based on the idea of h-likelihood, a generalization of theclassical likelihood to accommodate the random components coming through the model. It is preferable because itavoids the integration part for the marginal likelihood anduses the conditional distribution instead.Inspired by the idea of reinforcement learning and hierarchical models, this paper proposes a method by adding apenalty term to the h-likelihood. This method considers notonly the fixed effects but also the random effects in the linearmodel, and it produces good estimation results with theability to identify zero regression coefficients in joint modelsof mean-covariance structures for high-dimensional multilevel data.The rest of this paper is organized as follows: Section 2provides the literature review on current variable selectionmethods based on partial linear models and h-likelihood.Section 3 explains a penalty-based h-likelihood variable

2selection algorithm and demonstrates via simulation thatour proposed algorithm exhibits desired sample propertiesand can be useful in practical applications. Finally, Section 4concludes the paper, and some future research directions aregiven.2. Literature Review2.1. Reinforcement Learning in the Perspective of NonlinearSystems. Reinforcement learning, one of the most activeresearch areas in artificial intelligence, is introduced anddefined as a computational approach to learning wherebyan agent tries to maximize the total amount of reward itreceives when interacting with a complex, uncertain environment in Sutton and Barto [1]. In addition, in thepaper of Sutton and Barto [5], reinforcement learning isspecified to be trial and error (variation and selection andsearch) plus learning (association and memory). Furthermore, Barto and Mahadevan [6] propose hierarchicalcontrol architectures and associated learning algorithms.Approaches to temporal abstraction and hierarchicalorganization, which mainly rely on the theory of semiMarkov decision processes, are reviewed and discussed inBarto and Mahadevan’s paper [6]. Recent works, such asDietterich [7], have focused on the hierarchical methodsthat incorporate subroutines and state abstractions, instead of solving “flat” problem spaces.Nonlinear control design has gained a lot of attentionin the research area for a long time. In the industrial field,the controlled system usually has great nonlinearity.Various adaptive optimal control models have been applied to the identification of nonlinear systems in the pastliterature. In fact, the two important fundamental principles of controller design are optimality and veracity. Heet al. [8] study a novel policy iterative scheme for thedesign of online H optimal laws for a class of nonlinearsystems and establishes the convergence of the novelpolicy iterative scheme to the optimal control law. He et al.[9] investigate an online adaptive optimal control problem of a class of continuous-time Markov jump linearsystems (MJLSs) by using a parallel reinforcementlearning (RL) algorithm with completely unknown dynamics. A novel parallel RL algorithm is proposed, and theconvergence of the proposed algorithm is shown. Wanget al. [10] study a new online adaptive optimal controllerdesign scheme for a class of nonlinear systems with inputtime delays. An online policy iteration algorithm isproposed, and the effectiveness of the proposed method isverified. He et al. [11] propose the online adaptive optimalcontroller design for a class of nonlinear systems througha novel policy iteration (PI) algorithm. Cheng et al. [12]investigate the observer-based asynchronous fault detection problem for a class of nonlinear Markov jumpingsystems and introduces a hidden Markov model to ensurethat the observer modes run synchronously with thesystem modes. Cheng et al. [13] propose the finite-timeasynchronous output feedback control scheme for a classof Markov jump systems subject to external disturbancesand nonlinearities.Complexity2.2. Partial Linear Models. Linear models have been widelyused and employed in the literature. One extension of linearmodels, which was introduced by Nelder and Wedderburn[14], is generalized linear models (GLMs). GLMs allow theclass of distributions to be expanded from the normaldistribution to that of one-parameter exponential families.In addition, GLMs generalize linear regression in the following two manners: first of all, GLMs allow the linearmodel to be related to the response variable via a linkfunction, or equivalently a monotonic transform of themean, rather than the mean itself. Second, GLMs allow themagnitude of the variance of each measurement to be afunction of its predicted value.On the contrary, Laird and Ware [15] propose linearmixed effect models (LMEs), which are widely used in theanalysis of longitudinal and repeated measurement data.Linear mixed effect models have gained popular attentionsince they take into consideration within-cluster and between-cluster variations simultaneously. Vonesh andChinchilli [16] have investigated and applied statistical estimation as well as inference for this class of LME models.However, it seems that model selection problem in LMEmodels is ignored. This disregarded problem was noticedand pointed out by Vaida and Blanchard [17], stating thatwhen the focus is on clusters instead of population, thetraditional selection criteria such as AIC and BIC are notappropriate. In the paper of Vaida and Blanchard [17], theconditional AIC is proposed, for mixed effects models withdetailed discussion on how to define degrees of freedom inthe presence of random effects. Furthermore, Pu and Niu[18] study the asymptotic behavior of the proposed generalized information criterion method for selecting fixed effects. In addition, Rajaram and Castellani [19] use ordinarydifferential equations and the linear advection partial differential equations (PDEs) and introduce a case-baseddensity approach to modeling big data longitudinally.Recently, Fan and Li [20] develop a class of variableselection procedures for both fixed effects and random effects in linear mixed effect models by incorporating thepenalized profile likelihood method. By this regularizationmethod, both fixed effects and random effects can be selectedand estimated. There are two outstanding aspects regardingFan and Li’s [20] method. First of all, the proposed procedures can estimate the fixed effects and random effects in aseparate way. Or in other words, the fixed effects can beestimated without the random effects being estimated, andvice versa. In addition, the method works in the high-dimensional setting by allowing dimension of random effect togrow exponentially with sample size.Combined with the idea of generalized linear models(GLMs) and linear mixed effect (LME) models, one extension, generalized linear mixed models (GLMMs), is developed. In the traditional GLMs, it is assumed that theobservations are uncorrelated. To solve the constrainedassumption, GLMMs allow for correlation between observations, which often happens in the longitudinal data andclustered designs. The advantages of GLMMs are presentedas follows: first of all, GLMMs allow random effects to beincluded in the linear predictor. As a result, the correlations

Complexity3between observations can be explained through an explicitprobability model. Second, when the focus is on estimatingthe fixed effects on a particular individual, GLMMs providegood subject-specific parameter estimates. However, sinceGLMMs are also called multilevel models, it is generallymore computationally intensive when fitting the model.So far, all those GLMs and GLMMs are well-establishedparametric regression models. A serious disadvantage ofparametric modeling is that a parametric model may be toorestrictive in some applications. To overcome this restrictiveassumption difficulty in the parametric regression, nonparametric regression has gained popular attention in theliterature. There are many nonparametric and smoothingmethods, such as kernel smoothing, local polynomial fitting,and penalized splines. In this section, two often-usedsmoothing methods in estimating a nonparametric modelare described in the following paragraphs since they are usedlater in simulations and applications.The first type is called local linear kernel smoothing. Themain idea of local linear kernel smoothing is to locallyapproximate the function f linearly. Local linear kernelsmoothing uses Taylor expansion as a fundamental tool. Inparticular, Taylor expansion states that any smooth functioncan be locally approximated by a polynomial of some degree.Suppose we have a simple nonparametric modely i f t i εi ,(1)for i 1, . . . , n. Let t0 be an arbitrary fixed point where thefunction f is estimated. Assume f(t) has a first-ordercontinuous derivative at t0 . Then, by Taylor expansion, f(t)can be locally approximated byf(t) f t0 t t0 f(1) t0 ,(2)in a neighborhood of t0 that allows the above expansionwhere f(1) (t0 ) denotes the first derivative of f(t) at t0 .Let α0 f(t0 ) and α1 f(1) (t0 ). The local linearsmoother is obtained by fitting a data set locally with a linearfunction, to minimize the following weighted least squarescriterion:n2 yi α0 α1 t t0 Kh ti t0 ,(3)i 1where Kh (.) K(./h)/h, which is obtained by rescaling akernel function K(.) with a positive constant bandwidth h.The primary objective of the bandwidth h is to specify thesize of the local neighborhood [t0 h, t0 h], where thelocal fitting is conducted. Moreover, the kernel function K(.)determines how observations within the neighborhoodcontribute to the fit at t0 . A detailed introduction of thekernel function will be provided in the later paragraphs.The local linear smoother f h (t0 ) α 0 can be simplyexpressed asn i 1 s2 t0 s1 t0 t t0 Kh ti t0 yif , (4)h t0 s2 t0 s0 t0 s21 t0 wherens0 t0 Kh ti t0 ,i 1ns1 t0 Kh ti t0 ti t0 ,(5)i 1n2s2 t0 Kh ti t0 ti t0 .i 1A local linear smoother is often good enough for mostproblems if the kernel function K(.) and the bandwidth h areadequately determined. Moreover, it enjoys many goodproperties that the other linear smoothers may lack. Fan[21], Fan and Gijbels [22], and Hastie and Loader [23]separately discussed those good properties in detail.The kernel function K(.) used in the local linearsmoother is a symmetric probability density function. Thekernel K(.) specifies how the observations contribute to thelocal linear kernel fit at t0 , whereas the bandwidth h specifiesthe size of the local neighborhood [t0 h, t0 h]. Severalwidely used kernel functions include the following:(i)(ii)(iii)(iv)Uniform K(u) (1/2)I{ u 1}Epanechnikov K(u) (3/4)(1 u2 )I{ u 1}Biweight K(u) (15/16)(1 u2 )2 I{ u 1} 2Gaussian K(u) (1/ 2ϕ )e (1/2)uSuppose, for instance, the uniform kernel is used. All theti ’s within the neighborhood [t0 h, t0 h] contributeequally; or equivalently, the weights are the same, in the locallinear kernel fit at t0 ; on the contrary, all the ti ’s outside theneighborhood [t0 h, t0 h] contribute nothing. Suppose,for another example, the Gaussian kernel is used. Thecontribution of the ti ’s is determined by the distance of tifrom t0 . In other words, smaller distance (t t0 ) results inlarger contribution since the Gaussian kernel is a bell-shapedcurve, which peaks at the origin.The second type of smoothing is called regression splinesmoothing. In local linear kernel smoothing introducedabove, local neighborhoods were defined by a bandwidth hand a fixed point t0 . On the contrary, in regression splinesmoothing that will be introduced shortly, local neighborhoods are defined by a group of locations, known as knots,for example,τ 0 , τ 1 , . . . , τ K , τ K 1 ,(6)inaninterval[a, b],wherea τ 0 τ 1 · · · τ k τ k 1 b. Moreover, τ i , i 1, 2, . . . , kare referred as interior knots or simple knots. Then, localneighborhoods are divided by these knots, i.e., τ i , τ i 1 ,i 0, 1, . . . , k,(7)and within any two neighboring knots, a Taylor’s expansionup to some degree is applicable.A regression spline can be constructed in terms oftruncated power basis. As mentioned earlier, there are Kknots τ 1 , . . . , τ K , and the k-th degree truncated power basiscan be expressed as

4Complexitykk1, t, . . . , tk , t τ 1 , . . . , t τ K ,(8)where ak denotes power k of the positive part of a witha max(0, a). In most of the literature, it is called “constant, linear, quadratic, and cubic” truncated power basiswhen k 0, 1, 2, and 3 correspondingly. For the purpose ofthis chapter, cubic truncated power basis is used in subsequent sections of simulations and applications.We still consider the abovementioned simple nonparametric model:y i f t i εi ,(9)for i 1, . . . , n. It is with conventional purpose to denote thetruncated basis askk TΦp (t) 1, t, . . . , tk , t τ 1 , . . . , t τ K ,(10)where p K k 1 is the number of the basis functionsinvolved. Then, the regression fit of the function f(t) in thenonparametric model can be expressed as (t) Φ (t)T XT X 1 XT y,fpp(11)where y (y1 , . . . , yn )T and X (Φp (t1 ), . . . , Φp (tn ))T .To sum up, parametric models are very useful for longitudinal data analysis since they provide a clear and easydescription of the relationship between the response variableand its covariates. However, in most of data analysis, theparametric model does not fit the data well, resulting inbiased estimates. To overcome the restricted assumptions onparametric forms, various nonparametric models such asnonparametric mixed effects models have been proposed forlongitudinal data. Refer, for example, the study by Fan andZhang [24] and Wu and Rice [25] among others. There isalways a trade-off model assumption and model complexity.Parametric models are less robust against model assumptions, but they are efficient when the models are correctedassigned. On the contrary, nonparametric models are morerobust against model assumptions, but they are less efficientand more complex. A trade-off between efficiency andcomplexity by the information measure is fully investigatedand discussed in Caves and Schack [26]. Zhang et al. [27]propose an improved K-means clustering algorithm, whichis called the covering K-means algorithm (C-K-means).There are two advantages for the C-K-means algorithm. Firstof all, it acquires efficient and accurate clustering resultsunder both sequential and parallel conditions. Furthermore,it self-adaptively provides a reasonable number of clustersbased on the data features.Semiparametric models come across in the need tocompromise and remain good features of both parametricand nonparametric models. In semiparametric models,parametric component and nonparametric component arethe two essential components. More specifically, the parametric component is often used to model important factorsthat affect the responses parametrically, whereas the nonparametric component is often used for less important andnuisance factors. Various semiparametric models forlongitudinal data include semiparametric population meanmodels proposed in Martinussen and Scheike [28] and Xu[29], among others, and semiparametric mixed effectsmodels in the study by Zeger and Diggle [30], Groll and Tutz[31], and Heckman et al. [32]. For the purpose of this paper,we restrict our attention to partially linear regressionmodels.2.3. h-Likelihood. In longitudinal studies, there are two typesof models, marginal models, and conditional models. Bydefinition, marginal models are usually referred as population-average models by ignoring the cluster randomeffects. In contrast, conditional models have random effector are subject-specific models. The main difference betweenmarginal and conditional models is whether the regressioncoefficients describe an individual’s response or the marginalresponse to changing covariates. Or in other words,changing covariates does not attempt to control for unobserved subjects’ random effects. Diggle et al. [33] suggestedthe random effect model for inferences about individualresponses and the marginal model for inferences aboutmargins.The idea of h-likelihood was introduced by Lee andNelder [4]. h-likelihood is an extension of Fisher likelihoodto models of GLMs with additional random effects in thelinear predictor. The concept of h-likelihood is for inferencesof unobserved random variables. In fact, h-likelihood is aspecial kind of extended likelihood, where the random effectparameter is specified to satisfy certain conditions as we shalltalk more in details later. In the meantime, with the idea ofh-likelihood, hierarchical generalized linear models(HGLMs) were introduced as well in Lee and Nelder’s [4]paper. This class of hierarchical GLMs allows various distributions of the random component. In addition, thesedistributions are conjugate to the distributions of the response y. Four conjugate HGLMs were introduced in [4],namely, normal-normal, Poisson-gamma, binomial-beta,and gamma-inverse gamma (Table 1). If we let y be theresponse and u be the unobserved random component, v isthe scale on which the random effect u happens linearly inthe linear predictor. In other words, u and v are linked viasome strictly monotonic function.Consider the hierarchical model where y v and v followsome arbitrary distributions listed in Table 1. The definitionof h-likelihood, denoted by lh , is presented in the followingway:lh l(β, ϕ; y v) l(α; v),(12)where l(α; v) is the log likelihood function of v given parameter α and l(β, ϕ; y v) is that of y v given parameter βand ϕ. One point to note is that the h-likelihood is not atraditionally defined likelihood since v are not directlyobservable. In the traditional standard maximum likelihoodestimation for models with random effects, the method isbased on the marginal likelihood as the objective function. Inthis marginal likelihood approach, random effects v areintegrated out and what remain in the maximized functionare the fixed effects β and dispersion parameter ϕ. There are

Complexity5Table 1: Conjugate HGLMs.y uNormalPoissonBinomialGammauNormalGammaBetaInverse gammaLinkIdentityLogLogitLogtwo disadvantages of the marginal likelihood approach. Firstof all, the intractable integration of v is with obvious difficulty. In addition, random effects are nonestimable afterintegration. In contrast, the h-likelihood approach avoidssuch intractable integration. In fact, as clearly stated by Leeand Nelder [4], “we can treat the h-likelihood as if it were anorthodox likelihood for the fixed effects β and random effectsv, where the v are regarded as fixed parameters for realizedbut unobservable values of the random effects.” Furthermore, the h-likelihood allows us to have a fixed effect estimator that is asymptotically efficient as the marginalmaximum likelihood estimator. Last but not least, themaximized h-likelihood estimates are derived by solving thetwo equations simultaneously:zlh 0;zβ(13)zlh 0.zvPeople always expect an outstanding property of likelihood inference to be invariant with respect to transformations. As for maximum h-likelihood estimates, estimatesfor random effects are invariant with respect to the transformation of the random components of u.Furthermore, Lee and Nelder [4] mentioned adjustedprofile h-likelihood, which is defined in the following way:l(β) lh 1D lh log det ,2π2v v(14)where D(lh ) z2 lh /zv zvT . It eliminates the nuisance effects v from the h-likelihood. Moreover, the D(lh ) part isoften referred as the adjusted term for such elimination. Infact, this adjusted profile h-likelihood, which is used for theestimation of dispersion components, acts as an approximation of the marginal likelihood, without integrating v out.There are a few outstanding contributions in Lee andNelder’s [4] publication. First of all, it widens the choice ofrandom effect distributions in mixed generalized linearmodels. In addition, it brings about the h-likelihood as adevice for estimation and prediction in hierarchical generalized linear models. Compared to the traditional marginallikelihood, the h-likelihood avoids the messy integration forthe random effects and hence is convenient to use. Furthermore, maximized h-likelihood estimates are obtained byiteratively solving equation (14). To conclude, the h-likelihood is used for inference about the fixed and random effectsgiven dispersion parameter ϕ.On the contrary, Lee and Nelder [34] demonstrated theuse of an adjusted profile h-likelihood for inference aboutthe dispersion components given fixed and random effects.In this paper, the focus is on the joint modeling of the meanand dispersion structure. Iterative weighted least squares(IWLS) algorithm is used for estimations of both the fixedand random effects by the extended likelihood and dispersion parameters by the adjusted profile likelihood. Later,in [35], the algorithm was adjusted by replacing the extendedlikelihood to the first-order adjusted profile likelihood, as toestimate fixed effects in the mean structure.Lee and Nelder [36] proposed a class of double hierarchical generalized linear models in which random effectscan be specified for both the mean and dispersion. Compared with HGLMs, double hierarchical generalized linearmodels allow heavy-tailed distributions to be present in themodel. Random effects are introduced in the dispersionmodel to solve heteroscedasticity between clusters. Then,h-likelihood is applied for statistical references and efficientalgorithm, as the synthesis of the inferential tool. In addition,Lee and Noh [37] proposed a class of double hierarchicalgeneralized linear models in which random effects can bespecified for both the mean and dispersion, allowing modelswith heavy-tailed distributions and providing robust estimation against outliers. Greenlaw and Kantabutra [38]address the parallel complexity of hierarchical clustering.Instead of the traditional sequential algorithms, the described top-down algorithm in Greenlaw and Kantabutra[38] is parallelized and the computational cost of the topdown algorithm is with O(log n) time.In conclusion, for both hierarchical generalized linearmodels (HGLMs) and double hierarchical generalized linearmodels (DHGLMs), h-likelihood plays an important role ininferences for models having unobservable or unobservedrandom variables. Furthermore, numerical studies havebeen investigated and shown that h-likelihood gives statistically efficient estimates for HGLMs as well as DHGLMs.In addition, Noh and Lee [39] have shown that theh-likelihood procedure outperforms existing methods, including MCMC-type methods, in terms of bias. Last but notleast, compared to the traditional marginal likelihood, theh-likelihood avoids the messy integration for the randomeffects and hence is convenient to use. Therefore, theh-likelihood method is worth attention.3. Variable Selection via Penalized h-Likelihood3.1. Model Setup. Suppose that we have k independentgroups and each group contains m subjects. Let yij be the jthsubject of group i, where i 1, . . . , k and j 1, . . . , m. Basedon the idea of modeling the mean structure in the HGLMframework, we consider a partial linear model for modelingthe conditional mean:g μij f tij xTij β vi ,(15)where f(.) is an unknown smooth function in t, tij is anunivariate explanatory variable in [0, 1] for simplicity, g(.) isthe canonical link function for the conditional distribution

6Complexityof yij , and xij is a p 1 covariate vector with β as the associated coefficients. In matrix representation,y f(t) Xβ Zv ε.(16)We assume that conditional random variables ui and yijare from an exponential family with mean and variance: E yij ui μij ,(17) V yij ui ϕV μij .We also assume that (XT , t)T and ε are independent. Therandom effects presented in the mean model vi are linked toui via the relationship vi v(ui ), where ui N(0, σ 2u ). Thisallows for the definition of h-likelihood given in Lee andNelder [4]. In this paper, the identity link vi ui is used, andhence, this canonical scale corresponds to the case that theconditional distribution of the response y is normal, i.e.,yij N(μij , ϕ).For simplicity, random effects are considered in the formof a random intercept throughout this paper. If a randomintercept is not sufficient to represent the variation exhibitedin the data, then the model can be easily extended to a moregeneral form by considering a more complex random effectsstructure.3.2. Estimation Procedure via Penalized h-Likelihoodkm h likelihood f vi f yij vi i 1j 122m yij xTij β vi f tij 1vi 0 1 exp exp 222πσ2πui 1j 1k2km yij xTij β vi f tij 11v2i exp exp 2 kkm 22σ u j 12π σ u ( 2π ) i 1k2m(18)k2T2 σ u i 1 j 1 yij xij β vi f tij i 1 vi 11. km exp 2 k 2σ u2π σ u ( 2π )Thus, the log of h-likelihood is lh (β, v) lh (β, v) k log 2π σ u ( 2π )m 2k m i 1 j 1 yij xTij β vi f tij 22k i 1 v2i2σ 2u k log 2π σ u ( 2π )m lh (β, v) For the purpose of this paper, the first and second derivatives of lh (β, v) with respect to β and v are derived andlisted below:zβ zβT XT X;(20)zv zv(19)z2 lh (β, v)zlh (β, v)1 ZT (y Xβ Zv f(t)) 2 v;zvσuz2 lh (β, v)11‖v‖22 . ‖y Xβ Zv f(t)‖22 22σ 2uzlh (β, v) XT (y Xβ Zv f(t));zβT ZT Z 1I.σ 2uThe maximum likelihood estimate for the random effects v is obtained by setting zlh (β, v)/zv to zero. Then, an approximated likelihood for the fixed effects can be obtained byplugging the estimate v in lh (β, v). In addition, the marginallikelihood is approximated by the adjusted profile likelihood:

Complexity7l(β) lh (β, v) 1D lh (β, v) log det ,22πv vT(21)where D(lh (β, v)) z2 lh (β, v)/zv zvT .Now the problem of how to estimate the smoothfunction f(t) rises. In this paper, we use two nonparametricapproaches to estimate f(t): local linear regression technique and spline technique.In the framework of penalized variable selection, weapply a penalty on the approximated marginal likelihood sothatp lp (β) l(β) n Pλ βj ,(22)j 1where Pλ (.) is the penalty function with tuning parameter λ.Our aim is to maximize lp (β) and get the maximum likelihood estimates for the fixed effects β. We will give a brieftheoretical support on how to derive the estimation in thefollowing paragraphs.First of all, the L1 penalty functions are singular at theorigin, and they do not have continuous second-order derivatives. However, they can be locally approximated by aquadratic function as follows. Assume that

parametric regression models. A serious disadvantage of parametric modeling is that a parametric model may be too ctive assumption difficulty in the parametric regression, non-parametric regression has gained popular attention in the literature. ere are many nonparametric and smoothing

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3.4.2 Statically indeterminate structures 35 3.5 Element design 38 v. 3.5.1 General comments 38 3.5.2 Ties and struts 39 3.5.3 Beams and girders 40 3.5.4 Beam-columns 41 3.5.5 Members in portal frames 42 3.6 Examples 43 3.6.1 Ribbed dome structure 43 3.6.2 Two pinned portal—plastic design 45 .

Service Manual for WP6 Diesel Engine Preface WP6 mechanical pump series diesel engine has the features of compact structure, reliable

and disc/drum. While brake moan is still sensitive to the modal behavior of knuckle and sus-pension arms, a high frequency squeal in most cases exclusively depends on the properties of caliper, brake pad, and disc. While at high frequencies the distribution of energy via structure-borne sound is strongly

A. DREPT CIVIL GENERAL . I /1. PARTEA GENERALĂ. 1. Noţiunea actului juridic. Clasificarea actelor juridice după numărul părţilor, după scopul urmărit la încheierea lor, după efectul lor, după importanţa (gravitatea lor) şi după conţinutul lor. Bibliografie: 1) Gh. Beleiu, Drept civil român.