Unlike a pedigree initiated with two inbred lines, a full-sib family derived from two outbred parents frequently has many different segregation types of markers whose linkage phases are not known prior to linkage analysis.We formulate a general model of simultaneously estimating linkage, parental diplotype and gene order through multi-point analysis in a full-sib family. Our model is based on a multinomial mixture model taking into account different diplotypes and gene orders, weighted by their corresponding occurring probabilities. The EM algorithm is implemented to provide the maximum likelihood estimates of the linkage, parental diplotype and gene order over any type of markers.Through simulation studies, this model is found to be more computationally efficient compared with existing models for linkage mapping. We discuss the extension of the model and its implications for genome mapping in outcrossing species. The construction of genetic linkage maps based on molecular markers has become a routine tool for comparative studies of genome structure and organization and the identification of loci affecting complex traits in different organisms . Statist2 or backcross, always has two alleles and a consistent segregation ratio across different markers. Second, linkage phases among different markers are not known a priori for outbred parents and, therefore, an algorithm should be developed to characterize a most likely linkage phase for linkage analysis.Genetic analyses and statistical methods in outcrossing species are far more complicated than in species that can be selfed to produce inbred lines. There are two reasons for this. First, the number of marker alleles and the segregation pattern of marker genotypes may vary from locus to locus in outcrossing species, whereas an inbred line-initiated segregating population, such as an FTo overcome these problems of linkage analysis in outcrossoing species, Grattapaglia and Sederoff  proposedIn this article, we construct a unifying likelihood analysis to simultaneously estimate linkage, linkage phases and gene order for a group of markers that display all possible segregation patterns in a full-sib family derived from two outbred parents  is generated due to the combination of two-marker haplotypes [11] and [22], whereas diplotype [12] [21] (denoted by ) is generated due to the combination of two-marker haplotypes [12] and [21]. If the probability of forming diplotype [11] [22] is p, then the probability of forming diplotype [12] [21] is 1 - p. The genotype of parent Q and its possible diplotypes [33] [44] and [34] [43] can be defined analogously; the formation probabilities of the two diplotypes are q and 1 - q, respectively.In general, the genotypes of the two markers for the two parents can be observed in a molecular experiment, but the allelic arrangement of the two markers in the two homologous chromosomes of each parent (i.e., linkage phase) is not known. In the current genetic literatuire, a linear arrangement of nonalleles from different markers on the same chromosomal region is called the , 1 and , with a probability of pq, p(1 - q), (1 - p)q and (1 - p) (1 - q), respectively. The estimation of the recombination fraction in the full-sib family should be based on a correct diplotype combination [The cross of the two parents should be one and only one of four possible parental diplotype combinations, i.e., [11] [22] × [33] [44]), [11] [22] × [34] [43], [12] [21] × [33] [44] and [12] [21] × [34] [43], expressed as 11, 1bination . The foufor [11] [22] × [33] [44],for [11] [22] × [34] [43],for [12] [21] × [33] [44] and and , respectively.for [12] [21] × [34] [43]. Note that these matrices are expressed in terms of the combinations of the progeny genotypes for two markers n = (nj1j2)4 × 4 denote the matrix for the observations of progeny where j1,j2 = 1 for 13, 2 for 14, 3 for 23, or 4 for 34 for the progeny genotypes at these two markers. Under each parental diplotype combination, nj1j2 follows a multinomial distribution. The likelihoods for the four diplotype combinations are expressed asLet N1 = n11 + n22 + n33 + n44, N2 = n14 + n23 + n32 + n41, N3 = n12 + n21 + n34 + n43, and N4 = n13 + n31 + n24 + n42. It can be seen that the maximum likeihood estimate (MLE) of r  under the first diplotype combination is equal to one minus  under the fourth combination, and the same relation holds between the second and third diplotype combinations. Although there are identical plug-in likelihood values between the first and fourth combinatins as well as between the second and third combinations, one can still choose an appropriate  from these two pairs because one of them leads to  greater than 0.5. Traditional approaches for estimating the linkage and parental diplotypes are to estimate the recombination fractions and likelihood values under each of the four combinations and choose one legitimate estimate of r with a higher likelihood.where In this study, we incorporate the four parental diplotype combinations into the observed data likelihood, expressed asΘ = (r, p, q) is an unknown parameter vector, which can be estimated by differentiating the likelihood with respect to each unknown parameter, setting the derivatives equal to zero and solving the likelihood equations. This estimation procedure can be implemented with the EM algorithm [H be a mixture matrix of the genotype frequencies under the four parental diplotype combinations weighted by the occurring probabilities of the diplotype combinations, expressed aswhere lgorithm ,11,16. LwhereSimilar to the expression of the genotype frequencies as a mixture of the four diplotype combinations, the expected number of recombination events contained within each two-marker progeny genotype is the mixture of the four different diplotype combinations, i.e.,where the expected number of recombination events for each combination are expressed asDefineτ + 1}th EM step is given as follows:The general procedure underlying the {E Step: At step τ, using the matrix H based on the current estimate rτ}{, calculate the expected number of recombination events between two markers for each progeny genotype and ,dj1j2, hj1j2, pj1j2 and qj1j2 are the (j1j2)th element of matrix D, H, P and Q, respectively.where M Step: Calculate rτ+1} {using the equation,r converges to a value with satisfied precision. The converged values are regarded as the MLEs of Θ.The E step and M step among Eqs. (4) – (7) are repeated until A1, A2, A3 and A4, which are codominant to each other but dominant to the null allele, symbolized by O. Wu et al. [Unlike an inbred line cross, a full-sib family may have many different marker segregation types. We symbolize observed marker alleles in a full-sib family by u et al.  listed aA1A2 × A3A4, three non-null alleles A1A2 × A1A3, three non-null alleles and a null allele A1A2 × A3O, or two null alleles and two non-null alleles A1O × A2O;A. Loci that are heterozygous in both parents and segregate in a 1:1:1:1 ratio, involving either four alleles B. Loci that are heterozygous in both parents and segregate in a 1:2:1 ratio, which include three groups:1. One parent has two different dominant alleles and the other has one dominant allele and one null allele, e.g., A1A2 × A1O;B2. The reciprocal of B1;B3. Both parents have the same genotype of two codominant alleles, i.e., A1A2 × A1A2;BA1O × A1O;C. Loci that are heterozygous in both parents and segregate in a 3:1 ratio, i.e., D. Loci that are in the testcross configuration between the parents and segregate in a 1:1 ratio, which include two groups:1. Heterozygous in one parent and homozygous in the other, including three alleles A1A2 × A3A3, two alleles A1A2 × A1A1, A1A2 × OO and A2O × A1A1, and one allele  A1O × OO;D2. The reciprocals of D1.Dfully informative markers because of the complete distinction of the four progeny genotypes. The other six groups all contain the partially informative markers since some progeny genotype cannot be phenotypically separated from other genotypes. This incomplete distinction leads to the segregation ratios 1:2:1 (B), 3:1 (C) and 1:1 (D). Note that marker group D can be viewed as fully informative if we are only interested in the heterozygous parent.The marker group A is regarded as containing H for joint genotype frequencies between two fully informative markers. But for partially informative markers, only the joint phenotypes can be observed and, thus, the joint genotype frequencies, as shown in H, will be collapsed according to the same phenotype. Wu et al. [I) relating the genotype frequencies to the phenotype frequencies for different types of markers. Here, we use the notation  for a (b1 × b2) matrix of the phenotype frequencies between two partially informative markers, where b1 and b2 are the numbers of distinguishable phenotypes for markers  and , respectively. Correspondingly, we have . The EM algorithm can then be developed to estimate the recombination fraction between any two partial informative markers.In the preceding section, we defined a (4 × 4)-matrix u et al.  designedE Step: At step τ, based on the matrix (DH)' derived from the current estimate rτ}{, calculate the expected number of recombination events between the two markers for a given progeny genotype and :, ,  and  is the (j1j2)th element of matrices (DH)', H', P' and Q', respectively.where M Step: Calculate rτ+1} {using the equation,The E and M steps between Eqs. (8) – (11) are repeated until the estimate converges to a stable value.,  and . Let o1, o2 and o3 be the corresponding probabilities of occurrence of these orders in the parental genome. Without loss of generality, for a given order, the allelic arrangement of the first marker between the two homologous chromosomes can be fixed for a parent. Thus, the change of the allelic arrangements at the other two markers will lead to 2 × 2 = 4 parental diplotypes. The three-marker genotype of parent P (12/12/12) may have four possible diplotypes, [111] [222], [112] [221], [121] [212] and [122] [211]. Relative to the fixed allelic arrangement 1|2| of the first marker on the two homologous chromosomes 1 and 2, the probabilities of allelic arragments 1|2| and 2|1| are denoted as p1 and 1 - p1 for the second marker and as p2 and 1 - p2 for the third marker, respectively. Assuming that allelic arrangements are independent between the second and third marker, the probabilities of these four three-marker diplotypes can be described by p1p2, p1(1 - p2), (1 - p1)p2 and (1 - p1) (1 - p2), respectively. The four diplotypes of parent Q can also be constructed, whose probabilities are defined as q1q2, q1(1 - q2), (1 - q1)q2 and (1 - q1) (1 - q2) respectively. Thus, there are 4 × 4 = 16 possible diplotype combinations (whose probabilities are the product of the corresponding diplotype probabilities) when parents P and Q are crossed.Consider three markers in a linkage group that have three possible orders r12 denote the recombination fraction between markers  and , with r23 and r13defined similarly. These recombination fractions are associated with the probabilities with which a crossover occurs between markers  and  and between markers  and . The event that a crossover or no crossover occurs in each interval is denoted by D11 and D00, respectively, whereas the events that a crossover occurs only in the first interval or in the second interval is denoted by D10 and D01, respectively. The probabilities of these events are denoted by d00, d01, d10and d11, respectively, whose sum equals 1. According to the definition of recombination fraction as the probability of a crossover between a pair of loci, we have r12 = d10 + d11, r23 = d01 + d11 and r13 = d01 + d10. These relationships have been used by Haldane [Let  Haldane  to deriv), each corresponding to a diplotype combination, denoted by , where  for 1|2| or 2 for 2|1| denote the two alternative allelic arrangements of the second and third marker, respectively, for parent P, and  for 1|2| or 2 for 2|1| denote the two alternative allelic arrangements of the second and third marker, respectively, for parent Q. According to Ridout et al. [ are expressed in terms of d00, d01, d10 and d11.For a three-point analysis, there are a total of 16 (16 × 4)-matrices for genotype frequencies under a given marker order -matrices for the expected numbers of crossover that have occurred for The joint genotype frequencies of the three markers can be viewed as a mixture of 16 diplotype combinations and three orders, weighted by their occurring probabilities, and is expressed asSimilarly, the expected number of recombination events contained within a progeny genotype is the mixture of the different diplotype and order combinations, expressed as:Also defineThe occurring probabilities of the three marker orders are the mixture of all diplotype combinations, expressed, in matrix notation, asτ + 1}th EM step are given as follows:We implement the EM algorithm to estimate the MLEs of the recombination fractions between the three markers. The general equations formulating the iteration of the {E Step: As step τ, calculate the expected number of recombination events associated with D00(α), D01 (β), D10(γ), D11(δ) for the (j1j2j3)th progeny genotype :, , ,  and , (k = 1,2,3) usingCalculate nj1j2j3 denote the number of progeny with a particular three-marker genotype, hj1j2j3, , , , , pj1j2j3)1(, pj1j2j3)2(, qj1j2j3) 1(and qj1j2j3) 2(are the (j1j2j3)th element of matrices H, D00, D01, D10, D11, P1, P2, Q1 and Q2, respectively.where M Step: Calculate , ,  and  using the equations,d00, d01, d10 and d11 converge to values with satisfied precision. From the MLEs of the g's, the MLEs of recombination fractions r12, r13 and r23 can be obtained according to the invariance property of the MLEs.The E and M steps are repeated among Eqs. (19) – (32) until b1, b2 and b3, respectively. Define  is a (b1b2 × b3) matrix of genotype frequencies for three partially informative markers. Similarly, we define , and .Consider three partially informative markers with the numbers of distinguishable pheno-types denoted by Using the procedure described in Section (2.2), we implement the EM algorithm to estimate the MLEs of the recombination fractions among the three partially informative markers.m ordered markers on a linkage group. The joint genotype probabilities of the m markers form a (4m-1 × 4)-dimensional matrix. There are 2m-1 × 2m-1 such probability matrices each corresponding to a different parental diplotype combination. The reasonable estimates of the recombination fractions rely upon the characterization of a most likely parental diplotype combination based on the multilocus likelihood values calculated.Three-point analysis considering the dependence of recombination events among different marker intervals can be extended to perform the linkage analysis of an arbitrary number of markers. Suppose there are m-marker joint genotype probabilities can be expressed as a function of the probability of whether or not there is a crossover occurring between two adjacent markers,  where l1, l2, ..., lm-1 are the indicator variables denoting the crossover event between markers  and , markers  and , ..., and markers  and , respectively. An indicator is defined as 1 if there is a crossover and 0 otherwise. Because each indicator can be taken as one or zero, there are a total of 2m-1 D's.The  can be estimated using the EM algorithm. In the E step, the expected number of interval specific crossovers is calculated (see Eqs. (19) – (22) for three-point analysis). In the M step, an explicit equation is used to estimate the probability . The MLEs of  are further used to estimate m(m - 1)/2 recombination fractions between all possible marker pairs. In m-point analysis, parental diplotypes and gene orders can be incorporated in the model.The occurring probability of interval-specific crossover r = 0.05 vs. 0.20) and different sample sizes (n = 100 vs. 200).Simulation studies are performed to investigate the statistical properties of our model for simultaneously estimating linkage, parental diplotype and gene order in a full-sib family derived from two outbred parents. Suppose there are five markers of a known order on a chromosome. These five markers are segregating differently in order, 1:1:1:1, 1:2:1, 3:1, 1:1 and 1:1:1:1. The diplotypes of the two parents for the five markers are given in Table p or q) of parental diplotype is close to 1 or 0 (Table  and  in this example), two sets of MLEs,  = 1,  = 0 and  = 0,  = 1, give an identical likelihood ratio test statistic. Thus, two-point analysis cannot specify parental diplotypes for symmetrical markers even when the two parents have different diplotypes.As expected, the estimation precision of the recombination fraction depends on the marker type, the degree of linkage and sample size. More informative markers, more tightly linked markers and larger sample sizes display greater estimation precision of linkage than less informative markers, less tightly linked markers and smaller sample sizes Tables  and 2. Tr = 0.20) between markers  and  from two-point analysis is 0.0848, whereas this value from a three-point analysis decreases to 0.0758 when combining fully informative marker  but increases to 0.0939 when combining partially informative marker . The three-point analysis can clearly determine the diplotypes of different parents as long as one of the three markers is asymmetrical. In our example, using either asymmetrical marker  or , the diplotypes of the two parents for two symmetrical markers ( and ) can be determined. Our model for three-point analysis can determine a most likely gene order. In the three-point analyses combining markers , markers  and marker , the MLEs of the probabilities of gene order are all almost equal to 1, suggesting that the estimated gene order is consistent with the order hypothesized.The estimation precision of linkage can be increased when a three-point analysis is performed Table , but thiaa] [oo] × [aa] [oo] (cis × cis) and (2) [ao] [oa] × [ao] [oa] (trans × trans). The MLE of the linkage under combination (2), in which two dominant alleles are in a repulsion phase, is not as precise as that under combination (1), in which two dominant non-alleles are in a coupling phase [cis × cis, cis × trans, trans × cis and trans × trans). The combinations, cis × cis and trans × trans, have the same likelihood value, with the MLE of one combination being equal to the subtraction of the MLE of the second combination from 1. The same relationship is true for cis × trans and trans × cis. A most likely phase combination is chosen corresponding to the largest likelihood and a legitimate MLE of the recombination fraction (r ≤ 0.5) [To demonstrate how our linkage analysis model is more advantageous over the existing models for a full-sib family population, we carry out a simulation study for linked dominant markers. In two-point analysis, two different parental diplotype combinations are assumed: (1) [ng phase . For a gr ≤ 0.5) .aa] [oo] × [aa] [oo], one can easily select cis × cis as the best estimation of phase combination because it corresponds to a larger likelihood and a smaller   is biased since it is far less than the value of 0.20 hypothesized. Our model gives the same estimation precision of the linkage for the data derived from [ao] [oa] × [ao] [oa] as obtained when the analysis is based on a correct diplotype combination (Table  =  = 0 ).For our data set simulated from [r  Table . Our modns Table . In this of r 0.093 is biaaaa] [ooo] × [aaa] [ooo] and (2) [aao] [ooa] × [aao] [ooa]. The traditional approach is to calculate the likelihood values under three possible gene orders and choose one of a maximum likelihood to estimate the linkage. Under combination (1), a most likely gene order can be well determined and, therefore, the recombination fractions between the three markers well estimated, because the likelihood value of the correct order is always larger than those of incorrect orders  [rs Table . Howeverer Table . Our modOur model is further used to perform joint analyses including more than three markers. When the number of markers increases, the number of parameters to be estimated will be exponentially increased. For four-point analysis, the speed of convergence was slow and the accuracy and precision of parameter estimation have been affected for a sample size of 200 (data not shown). According to our simulation experience, the improvement of more-than-three-point analysis can be made possible by increasing sample size or by using the estimates from two- or three-point analysis as initial values.AaVvXx for markers ,  and . Because these three markers are dominant, the cross generates 8 distinguishable genotypes, with observations of 28 for A-/V-/X-, 4 for A-/V-/xx, 12 for A-/vv/X-, 3 for A-/vv/xx, 1 for aa/V-/X-, 8 for aa/V-/xx, 2 for aa/vv/X- and 2 for aa/vv/xx. We first use two-point analysis to estimate the recombination fractions and parental diplotypes between all possible pairs of the three markers. The recombination fraction between markers  and  is , whose the estimated parental diplotypes are [Av] [aV] × [AV] [av] or [AV] [av] × [Av] [aV]. The other two recombination fractions and the corresponding parental displotypes are estimated as , [Vx] [vX] × [VX] [vx] or [VX] [vx] × [Vx] [vX] and , [AX] [ax] × [AX] [ax], respectively. From the two-point analysis, one of the two parents have dominant alleles from markers  and  are repulsed with the dominant alleles from marker .We use an example from published literature  to demon. The MLEs of the recombination fractions are ,  and . The parental diplotype combination is [XAV] [xav] × [XAv] [xaV] or [XAv] [xaV] × [XAV] [xav]. The three-point analysis for these three markers by Ridout et al. [ was not considered.Our subsequent three-point analysis combines parental diplotypes and gene orders to estimate the linkage between these three dominant markers. The estimated gene order is t et al.  led to tSeveral statistical methods and software packages have been developed for linkage analysis and map construction in experimental crosses and well-structured pedigrees -6, but tIn this article we present a unifying model for simultaneously estimating the linkage, parental diplotype and gene order in a full-sib family derived from two outbred parents. As demonstrated by simulation studies, our model is robust to different parameter space. Compared to the traditional approaches that calculate the likelihood values separately under all possible linkage phases or orders ,10,18, oThird, and most importantly, our unifying approach can significantly improve the estimation precision of the linkage for dominant markers whose alleles are in repulsion phase. Previous analyses have indicated that the estimate of the linkage between dominant markers in a repulsion phase is biased and imprecise, especially when the linkage is not strong and when sample size is small . There aOur approach will be broadly useful in genetic mapping of outcrossing species. In practice, a two-point analysis can first be performed to obtain the pairwise estimates of the recombination fractions and using this pairwise information markers are grouped based on the criteria of a maximum recombination fraction and minimum likelihood ratio test statistic . The parIn any case, our two- and three-point analysis has built a key stepping stone for map construction through two approaches. One is the least-squares method, as originally developed by Stam , that caOur maximum likelihood-based approach is implemented with the EM algorithm. We also incorporate the Gibbs sampler  into theQL derived the genetic and statistical models and wrote computer programs. YHC participated in the derivations of models and statistical analyses. RLW conceived of ideas and algorithms, and wrote the draft. All authors read and approved the final manuscript.


Longitudinally observed cohort data can be utilized to assess the potential for health promotion and healthcare planning by comparing the estimated risk factor trends of non-intervened with that of intervened. The paper seeks (1) to estimate a natural transition (patterns of movement between states) of health risk state from a Korean cohort data using a Markov model, (2) to derive an effective and necessary health promotion strategy for the population, and (3) to project a possible impact of an intervention program on health status.The observed transition of health risk states in a Korean employee cohort was utilized to estimate the natural flow of aggregated health risk states from eight health risk measures using Markov chain models. In addition, a reinforced transition was simulated, given that a health promotion program was implemented for the cohort, to project a possible impact on improvement of health status. An intervened risk transition was obtained based on age, gender, and baseline risk state, adjusted to match with the Korean cohort, from a simulated random sample of a US employee population, where a health intervention was in place.The estimated natural flow (non-intervened), following Markov chain order 2, showed a decrease in low risk state by 3.1 percentage points in the Korean population while the simulated reinforced transition (intervened) projected an increase in low risk state by 7.5 percentage points. Estimated transitions of risk states demonstrated the necessity of not only the risk reduction but also low risk maintenance.The frame work of Markov chain efficiently estimated the trend, and captured the tendency in the natural flow. Given only a minimally intense health promotion program, potential risk reduction and low risk maintenance was projected. Evidence was found that health promotion programs affect health risks in the US and in many other countries -4. Also,A previous study with a random sample from a 12-year cohort of civil servants in Korea  providedLongitudinally observed cohort data can be utilized for health promotion and healthcare planning provided the health risk trend is estimated and it poses a general need for an intervention. This paper attempts to address an issue around the possibility of predicting the impact of a health promotion intervention by applying observed effectiveness data from a population with an intervention to the observed transitions of risk status in a Korean population in the absence of such a program.To help in understanding the overall risk transitions in the Korean population and to implement an effective intervention program, a Markov chain model was utilized, assuming finite risk states at any point in time . Today'sThis paper utilizes the observed transitions of measured health risks in a cohort of the Korean National Health Insurance Corporation registrants (KNHIC) over 5 years. This longitudinally-followed population trend, without a particular intervention or policy change in place, was used as a basis to estimate a natural history of risk flow. In addition, an intervened transition was simulated, given that a health promotion program was implemented for the KNHIC cohort to project a possible impact on the improvement of health status. Comparing the two transitions also provided directions for a health promotion program that might be implemented in this population.The study population consisted of the established KNHIC prospective cohort . RegistrSimilarly, a comparison population was selected for the simulation of a program effect from the large longitudinal database of University of Michigan-Health Risk Appraisal (UM-HRA) completers. The UM-HRA was originally a CDC version, which was modified to fit the national trend of cost, and to meet the guidelines over time. Additional conditions applied are: (1) participated in a health promotion program at a minimal intensity, (2) completed the UM-HRA at least three times during 1996–2000, (3) were insured by the same insurance plans during the corresponding years and (4) were actively employed over the period at the same industry, and (5) age under 65 years in 1996 (N = 15,793).The survey questionnaires (p-HRA), on health status, diet, and lifestyles, were sent to work places and homes to encourage a national health screening at designated health care facilities and to measure lifestyle related health behaviors, every two years for the KNHIC registrants. UM-HRA was used to appraise individual health status during the same period 1996–2000) for the US population. The validity of UM-HRA has been addressed elsewhere [96–2000 fThree lifestyle-related health risks were measured by the corresponding questionnaires (p-HRA for the study cohort, and UM-HRA for the comparison population): physical activity, alcohol consumption, and smoking. In addition, pivotal measures of overall health were collected by the questionnaires: perceived health and medical conditions. Three biometric measures were obtained from the appropriate lab tests during physical examination for the KNHIC population and from the self-reported measures for the comparison population: systolic blood pressure (SBP) /diastolic blood pressure (DBP), total cholesterol, and body mass index (BMI) via height and weight measures. The risk criteria for the US population were defined Table  followinInformation on eight health risks for the study cohort were systematically evaluated and mapped to the measured risks by UM-HRA Table . This waInpatient plus outpatient costs per annum were collected from KNHIC database for the medical utilization in association with health risks, and the inflation adjusted average 1996 costs (January 1st, 1996 through December 31, 1996) were used for the T1 costs. Similarly, average T2 costs were calculated from 1998 claims costs.Participants of the p-HRA were not given any further information on identified health risks. Neither was it used to gain access to any health intervention programs during the five years (1996–2000). In US, on the other hand, as part of an intervention program, the completers of UM-HRA were given individually tailored health information, followed by encouragement of participation in a health promotion program at no cost during 1996–2000. This nation-wide program included an annual mail-based HRA, personalized follow-up report, identification of top significant risks and referral to health resources. This was defined as minimal level intervention, which differs from KNHIC's p-HRA in providing health information and individual feedback.rd time frame, 2000 for the KNHIC population, and 1998 or 2000 for the US population. Population with minimal intervention was matched to the baseline characteristics of KNHIC population, using age, gender, and baseline risk distribution. Trend was defined as the risk state change in each population between the time-points while each transition was annualized. Each change of risk state was estimated and the parameters to trend from the matched intervened population were used to project the possible trend of KNHIC population, following such a program.Provided that the least resources were available for a health intervention of the KNHIC population, a projected health risk transition with the minimal level intervention was utilized for the simulation of possible impact on risk transition. Population health trends were followed over the three time frames (T1, T2, and T3). T1 refers to the baseline year, which is 1996 for both populations. T2 refers to the second time frame, which is 1998 for the KNHIC population, and 1997 or 1998 for the compared US population. T3 refers to the 3An age/gender-cohort model was implemented, following observations on the natural flow of health status over three time frames, and it was compared to the corresponding age/gender-cohort of a US population with the matched risk distributions, where an intervention at minimal level was applied.For a simulation of an intervened transition for KNHIC population, an adjusted estimation of intervened transition was obtained based on random samples drawn from the selected US sample of intervened employees (N = 15, 793). To simulate additional random samples to reduce variances in the estimate of parameters for transition, a Monte Carlo bootstrap ,27 methoComputations were carried out using a generalized linear model for the ordered categorical outcomes (low risk<medium risk<high risk), assuming a Markov chain model in which the order of dependency was learnt from the data Figure . Figure Assumptions of Markov chain model, and model fit diagnostics were evaluated for the KNHIC population. Stationarity was tested by likelihood ratio test of the fitted models using a generalized equation estimation with and without time-dependence while controlling for other measured covariates. Order of Markov chain was tested with likelihood ratio test and fit-diagnostics were run. Following the fitted Markov model, we presented the estimated transitions over three time-frames for the natural flow. Cochran-Armitage trend tests controlling for T1 or T2 risk state were performed to test the trend of dependency of future state on T2 or T1 risk state.Average costs in each T1-T2 risk state pair were calculated from the corresponding years' average costs, adjusted for inflation. Healthcare costs were converted to ratios of average costs per T1-T2 risk transition state, compared to low (T1)-low (T2). Adjusting with any appropriate inflation rate in the future, these ratios can be used to project the cost savings over time by comparing the number of projected trends times cost ratios in natural and intervened flows. The projected percentage changes in the two trends  were compared for an estimation of an intervention impact on risk change.Individual baseline risk prevalence of the two cohorts is shown in Table Under the assumption of finite number of health risk states (low, medium and high states), Markovian models were examined for proper order estimation, an association structure, and stationarity. Following a likelihood ratio test, Markov chain (MC) order 2 was preferred for the risk transition of the non-intervened (KNHIC) population (Table When Markov chain order 2 was assumed, higher T1 risk status, controlling for T2 risk state as in Table Cochran-Armitage trend tests controlling for T1 risk state or T2 risk state show a strong declining trend of staying at "0" than that at T2 as in the order of appearance in Table We applied the results from the previous sections (regarding the KNHIC population) to the observed risk transitions of the matched sample from a US active employee population, who participated in a minimal level program. Similar assumptions were tested and stationary Markov chain order 2 was also applied. The estimated transition probabilities without an intervention (KNHIC cohort) and those of the same people with a minimal level intervention are shown in Table Overall, the likelihood of maintaining health at low risk (T1-T2-T3) is higher in the intervened transition (0.97 vs. 0.81). Also, the probabilities of being at high (T1-T2-T3) and at medium (T1-T2-T3) are lower in the intervened flow (0.54 vs. 0.59 at high, 0.32 vs. 0.44 at medium). Healthcare cost ratios in relation to the T1 -T2 risk states in the KNHIC population are also shown in the table. In general, cost ratios increased in the order of T1-T2 risk states and cost utilization of high-high group almost doubled (1.79) the cost of low-low.rd power of the Markov transition matrix of order 2. This is then collapsed by the baseline states as in Table After three years of projection, the transitions, presented in Table After controlling for age, gender and other covariates, such as past healthcare costs, variations within groups in predicting the future risk transition were left unexplained. This suggested an additional consideration of dependence on other factors such as past health risk history. This dependent structure of a natural risk flow was best fitted with a Markov model order 2 (due to limitation of the data, no higher order was tested). In other words, current risk states largely depend on the immediate past state and also depend on the one before that. The estimated probability was stable. There was no policy change or environmental effect, which is considered as a period effect, during the study period. In addition, as a result of testing a higher order MC for the available US data, order 2 was preferred to the order 3 (data not shown). Therefore, we concluded that an assumption of MC order 2 was plausible for the health risk transitions regardless of natural flow or reinforced flow at minimal intensity ,29. AlsoSimulated data for a minimal program participation, adjusted to matched age, gender, and baseline risk distribution of KNHIC population for comparison, showed that sampling disparities were significantly reduced in overall distribution and demographics but still remained in individual risk factors. Exercise and smoking risks are relevant to environmental and cultural adaptation, which showed large differences in prevalence. Typically, overweight is the most prevalent risk factor in the US, and it appeared to be significantly higher even after matching overall risk distribution to the Korean population. Although there exist some differences, these biases in the data were assumed to be smoothed out by collapsing to the aggregated risk levels.There may be inherent differences in risk transitions in the two populations due to two factors: (1) disparities in the individual risk factors, 2) systematic differences due to cultural and environmental factors. Due to the fact that estimation of risk transition was primarily dependent on the aggregated risk state rather than the individual risks  risk state, given age  of risk reduction program should differentiate this path-variability to maximize the impact Tables . FollowiOver the period, although there was presumably an aging process in the natural flow, there exists a regression to the previous state. Especially with Markov chain order 2, in controlling for the T1 state, the tendency stays. This implies that even without any external reinforcement for a healthier state, low risk state largely tends to maintain its state, followed by the high risk state. Therefore, an intervention with minimal effort to sustain healthy lifestyle and behavior for those who are at low risk may yield substantial benefit in the long run Tables ,4. LikewFor this study population, the minimal level intervention appeared to be effective in the low and medium baseline groups Table . This isOne of the study goals was to achieve estimate a natural flow of the study population, where a demand and necessity for health promotion program has been increased recently. Natural flow was coined by Edington  to descrIt is recommended to set a policy and allocate resources, tailored to the population profile of aggregated risk state, first. Targeting risk factors in the context of risk state or risk clusters could follow next. For example, an overweight population at high risk state requires more resources and more intensive interventions than the overweight population at low risk state. The study provides the base information when planning such a population-based intervention. This approach differentiates from those interventions targeting individual risk profiles, which could be inconsistent at times. Studies suggest that an inter-association over risks is more important in implementing an intervention than individual risks . Some vaAs a reasonable method to project the risk trends, a stationary Markov model of order 2 was fitted for the health risk transitions of the non-intervened population, suggesting a natural risk flow for the population. Utilizing this with the matched intervened population, the reinforced risk transitions of the KNHIC population were estimated. The significant difference in the transitions appeared to be for the low baseline risk population even with a minimal intensity intervention program. Therefore, the difference in the projected numbers by the two transitions (with /without interventions) showed a significant impact on low risk maintenance although higher risk population was also impacted to increase fewer risks and to moderate it to becoming "medium risk".Conventionally, most programs in the past have focused on risk reduction although low risk maintenance has been raised as a practical issue. Since this paper suggests a strong dependence on the previous history of risk status and the instability of high (T1)-low (T2) risk population, compared to low-to-low or medium-to-low population, differentiating efforts for the low risk people (from low to low) and for those moved to low (from high to low), should be sought for maximum impact.The studied intervention with minimal intensity was to diversify the beneficiaries of the program, to increase awareness, to continue educating and motivating individuals to adopt healthier behaviors by individualized feedback, and to provide resources . The finPopulation characteristics on risk transition were assumed to be similar by adjusting for age, gender, if previous risk states were the same. However, in addition to the measured risks, diet and culturally inherited behavioral differences could make an inherent transitional difference in the two compared populations. Also, some risks such as disease were underestimated in the KNHIC population. Therefore, the presented natural flow estimation in Korean population can be adopted but utilized with caution.It may be conjectured that there are some risk factors are more easily modifiable than others such as exercise while medical condition and overweight may not. The currently shown disparity in the two compared risk profiles may have induced somewhat less significant program impact on risk transition (lowering risk state) due to higher prevalence in medical conditions and overweight in the control population. This study tried to match two populations as close as possible and to model to reduce potential biases but comparing two populations requires more caution in further relevant studies.Despite the fact that individual risk distribution is not consistent often time, not only an overall health status but a multivariate risk state (i.e. a matrix per person) could be utilized to identify an effect of each risk factor to their projected state in future, as in Manton and Stallard . Based oThe authors declare that they have no competing interests.JP carried out the analyses and drafted the manuscript. SHJ collected, cleaned the data and provided initial analyses. DWE participated in the design of the analyses and provided revisions. All authors read and approved the final manuscript.


To adequately assess individual studies and synthesize quantitative research on weight loss studies, transparent reporting of data is required. The authors examined the reporting quality of randomized trials in the weight loss literature, focusing exclusively on subject characteristics as they relate to enrollment, allocation, and follow-up.An extensive literature review, which included a computerized search of the MEDLINE database, manual searches of bibliographic references, and cross-referencing of 92 review articles was conducted. A checklist, based on CONSORT recommendations, was used to collect information on whether or not authors reported age, gender, co-morbid disease, medication use, race/ethnicity, and postmenopausal status. Also tracked was whether or not initial and final sample size was reported and stratified by gender.Of 604 possible articles, 231 articles met eligibility criteria. Important subject characteristics were not reported as the following breakdown indicates: age (11%), gender (4%), race/ethnicity (86%), co-morbid disease states (34%), and medication use (92%). Additionally, 21% of articles failed to report initial sample size by gender while 69% neglected to report final sample size by gender.Inadequate reporting can create difficulties with interpretation and can lead to biased results receiving false credibility. The quality of reporting for weight loss studies needs considerable improvement. Current statistics indicate that approximately 64.5% of US adults can be considered overweight while 30.5% can be classified as obese . ObesitySystematic reviews can offer the most reliable sources of information on which to develop guidelines and base treatment policy. However, such reviews need to contain a high proportion of all relevant evidence, which relates both to the need to find all trials and the need to analyze data on all participants . To imprIn response to the need for quality reporting, a panel of clinical investigators, epidemiologists, biostatisticians, and journal editors published a statement called the Consolidation of the Standards of Reporting Trials (CONSORT), which is designed to improve the standard of written reports of RCTs . The CONFor the present study, we used information collected from a meta-analysis of the diet and exercise literature. The primary purpose of the meta-analysis was to statistically integrate and analyze published research studies on the effects of diet restriction only, diet restriction and exercise, or exercise only on weight loss, body composition, fat distribution, metabolism, and aerobic fitness. We examined the frequency of explicit reporting of subject characteristics (i.e., age, gender, co-morbid disease states, medication use (in addition to intervention drug), race/ethnicity, and postmenopausal status). We also examined whether or not initial and final sample size was reported and if so, we recorded if the final sample size was reported by gender.Search strategy. With the assistance of a clinical medical librarian, we developed search strategies, key words, and check tags to begin our initial search for relevant articles to include in the meta-analysis. We identified studies from a computerized search of the MEDLINE database  from 1966 – 2003, manual searches of bibliographic references of relevant published articles, and extensive cross-referencing and manual searches of 92 review articles.2) adult participants (≥ 18 years old). Ambulatory patients who were kept on wards exclusively for study purposes were also included. However, studies including pregnant women, patients with serious medical conditions, military personnel, and trained or professional athletes were excluded.Subject types. We required that intervention trials must study overweight or obese  active intervention trials with one group assigned to a weight-loss program involving energy restriction, exercise (aerobic or weight training) or both energy restriction and exercise; 2) outcome measures included weight loss; however, weight loss was not required to be a primary outcome; 3) the minimum duration of the intervention for weight loss must be greater than one day; and, 4) the article must be published in the English language. Case studies, crossover trials (due to possible carryover effects of weight reduction), and exploratory studies were excluded. We did not require studies to have been conducted in the United States. Pharmacotherapy, hormonal therapy, or surgical treatment studies, which did not include a separate group receiving a weight loss intervention (i.e., diet and/or exercise) were not included in the review process.Initial screening process. We independently screened all records resulting from our MEDLINE search strategies and manual searches by examining the titles and abstracts. We rejected articles if we could determine from the title and/or abstract that the study did not meet our selection criteria. If we could not determine whether or not to reject the article or if there was disagreement among reviewers, we retrieved the full text of the article for further evaluation.To control for abstraction bias, we trained five graduate research assistants to extract information from articles, using sample articles. Reliability checks were completed on the first several sample articles before graduate students were allowed to code independently. On a monthly basis thereafter, inter- and intra-rater reliability checks were completed to control for drift. All research assistants were required to attend weekly meetings with the principal investigator and co-investigators to discuss coding issues and to resolve coding disagreements.To control for potential selection biases after our initial screening procedure, we selected or rejected studies based on an examination of the research design and methodology. Acceptance or rejection for inclusion in the meta-analysis was not based on the study's outcome. We required coders to list specific reasons for all excluded studies. In addition, we tried to enhance data reporting for studies with missing covariate data by contacting authors by email or letter.Data entry quality control measures included the design of screens to match the coding sheet, automatic range limits placed on items entered, and queries to check for outliers. Values greater than two standard deviations from the mean were checked against the article. In addition, a random sampling of 58 articles (~25%) was pulled to check for accuracy.Standardized coding sheet for data abstraction. A separate checklist was devised based on CONSORT recommendations and quality indicators suggested by other authors -10. SpecThe frequency of reporting in the journal articles for the selected variables was obtained by performing a proc frequency using SAS (Version 8.2, Cary, NC). Results are presented as totals.Of the 604 randomized clinical trial articles that were reviewed, 231 (38%) met eligibility criteria for the present study. Figure American Journal of Clinical Nutrition contributed the most articles that met the eligibility criteria (47), followed by the International Journal of Obesity (32), Medicine & Science in Sports & Exercise (24), Journal of the American Medical Association (11), Metabolism (11), New England Journal of Medicine (7), Archives of Internal Medicine (6), and Journal of the American Dietetics Association (6). In addition, 52 journals had 5 or fewer articles in this analysis.For this analysis, the Of the 231 articles that met eligibility criteria for the present study, subjects' age was not reported in 25 (11%) articles. Investigators of ten (4%) different studies failed to report the gender of their study participants. Race and/or ethnicity of study subjects were not reported in 199 (86%) studies. In addition, 78 (34%) studies failed to report the health status (i.e., presence or absence of co-morbidities) of their study subjects at baseline. Further, medication use was not reported in 213 (92%) of the articles  studies while final sample size was not reported in 133 (58%) of the 231 eligible articles. When reporting initial and final sample size by gender, 68 articles were included in the analysis. Of the eligible articles, 14 (21%) did not report initial sample size by gender while 47 (69%) failed to report final sample size by gender  in the literature. It is well known that certain disease states, such as diabetes and cancer, affect metabolism -17, whicMenopause is associated with decreases in lean mass and increases in fat-mass -22. TherAttrition may threaten the internal and external validity of the scientific literature as well as the efficacy of a specific dietary/exercise intervention ,24. For In the present critical appraisal of the methodological quality of weight loss studies, we found major shortcomings in the reporting of subject characteristics as they relate to enrollment, allocation, and follow-up in trials that evaluated diet and exercise interventions. Many studies did not report variables that may explain some of the variance in outcomes. These findings are consistent with those of similar studies, which indicate inadequate reporting of subject characteristics ,25 and rConsolidation of the Standards of Reporting Trials (CONSORT), randomized controlled trials (RCTs), randomized trials (RTs), kilojoules per day (kJ/d), standard deviation (SD), kilograms (kg), years (yrs)The author(s) declare that they have no competing interests.CAG made substantial contributions to the conception and design, acquisition of data, analysis and interpretation of data, and drafting and revising the manuscript for important intellectual content. EPK, JDL, BWB and GH made substantial contributions to the submitted manuscript by assisting with the acquisition of data, analysis and interpretation of data, drafting of the manuscript, statistical analyses, and critical revisions. JED made substantial contributions to the manuscript by helping with the conception and design, critical revisions of the manuscript, and obtaining funding. All authors read and approved the final manuscript.Supported by National Institutes of Health NIH DK56303, Joseph E. Donnelly, Ed.D., principal investigatorThe pre-publication history for this paper can be accessed here:


There is little in biology that compares in beauty and limpidity to the development of a zebrafish embryo as viewed through a light microscope. The transparent eggshell and embryo tissues expose the minutest details of cell migrations and organ assembly to the curious viewer. Within a day, distinct vertebrate features emerge: a distinct head with the outlines of two large eyes, a quickly pumping heart, a notochord, and a growing array of somites—the bone and muscle precursors—stretching from trunk into tapering tail.The transparent zebrafish embryo has allowed geneticists to discover a large number of mutants with anomalies in the development of external and internal organs. Seven mutations, collectively known as “You-class,” turn the pointed, chevron-like somites into shallow, rounded arcs (“You” stands for “U-shaped”). Ian Woods and William Talbot now show that the You mutation disrupts a new modulator of Hedgehog signaling.Hedgehog is an extracellular signaling protein that can impose various fates on target cells at close proximity or over longer distances. Much research is focused on understanding the factors that promote or limit Hedgehog's activity and range. Woods and Talbot propose that the You protein acts in the extracellular environment to promote Hedgehog signaling.Hedgehog was originally named for mutations that cause excess brush-like denticles to grow on the surface of fruitfly embryos, but it is now known to direct countless developmental decisions in invertebrates and vertebrates alike. In addition, several cancers are known to result from inappropriate Hedgehog signaling. In fish, Hedgehog's best-documented role is in muscle development. In the absence of Hedgehog signaling, cells destined to become slow muscle fibers fail to differentiate properly. A subset of these slow muscle cells—the muscle pioneers—congregate near the dorso-ventral midline of the embryo, where the dorsal and ventral halves of somites converge. When these specialized cells are absent, abnormal somite assembly leads to the U-shaped phenotype.The authors found that You mutants showed many telltale signs of reduced Hedgehog signaling. Proteins that are normally expressed at certain times during the development of slow muscle cells were not activated in You mutants, indicating that these cells did not form. Mutant embryos also displayed reduced expression of the Hedgehog receptor Patched, a universal reporter of Hedgehog signaling activity. In addition, You mutants had specific ventral spinal chord defects that are shared by known Hedgehog pathway mutants. Yet You mutants expressed Hedgehog normally. Moreover, Hedgehog targets could still be activated in You mutants in response to excess Hedgehog signaling, suggesting that the signaling cascade is left intact. The authors concluded that the You protein was a facilitator rather than a crucial transmitter in Hedgehog signaling, likely acting at a step upstream of a cell's response to Hedgehog. Normal muscle pioneers could form in chimeric embryos (embryos made of wild-type and You mutant cells) regardless of which cells—the Hedgehog-producing cells or Hedgehog-responding muscle precursors—expressed You. This made it most likely that the You protein acted outside the cells, perhaps as a cell matrix component.The authors mapped the You mutation and found that it disrupted the coding region of a gene encoding a putative secreted protein. The predicted You protein is closely related to members of the mouse SCUBE family, a group of proteins that are defined by characteristic extracellular motifs . This observation strengthens the hypothesis that the You protein has extracellular functions, and the researchers' experimental evidence supports a role for You in transport or stabilization of Hedgehog. At later stages You could also participate in other signaling pathways, as its expression does not always coincide with that of Hedgehog during zebrafish development.


Intrathecal baclofen pump has been used effectively with increasing frequency in patients with severe spasticity, particularly for those patients who are unresponsive to conservative pharmacotherapy or develop intolerable side effects at therapeutic doses of oral baclofen. Drowsiness, nausea, headache, muscle weakness, light-headedness and return of pretreatment spasticity can be caused by intrathecal pump delivering an incorrect dose of baclofen. Intrathecal baclofen withdrawal syndrome is a very rare, potentially life-threatening complication of baclofen pump caused by an abrupt cessation of intrathecal baclofen.A 24-year-old man with a past medical history of cerebral palsy and spastic quadriparesis developed hyperthermia, disseminated intravascular coagulation, rhabdomyolysis, acute renal failure and multisystem organ failure leading to a full-blown intrathecal baclofen withdrawal syndrome. Intrathecal baclofen pump analysis revealed that it was stopped due to some programming error. He was treated effectively with supportive care, high-dose benzodiazepines and reinstitution of baclofen pump.The episodes of intrathecal baclofen withdrawal syndrome are mostly caused by preventable human errors or pump malfunction. Educating patients and their caregivers about the syndrome, and regular check-up of baclofen pump may decrease the incidence of intrathecal baclofen withdrawal syndrome. Oral baclofen replacement may not be an effective method to treat or prevent intrathecal baclofen withdrawal syndrome. Management includes an early recognition of syndrome, proper intensive care management, high-dose benzodiazepines and prompt analysis of intrathecal pump with reinstitution of baclofen. Baclofen is a gamma-aminobutyric acid (GABA) analog that has inhibitory effects on spinal cord reflexes and brain. Intrathecal baclofen (ITB) therapy consists of long-term delivery of baclofen to the intrathecal space. Intrathecal baclofen has been used to treat spasticity due to cerebral palsy, brain or spinal cord injury, multiple sclerosis, dystonia, stroke and stiff-man syndrome, particularly for those patients who are unresponsive to conservative pharmacotherapy or develop intolerable side effects at therapeutic doses of oral baclofen . Side efA 24-year-old man was admitted to our intensive care unit (ICU) with a possible diagnosis of seizure disorder and sepsis. He had a past medical history of cerebral palsy and spastic quadriparesis. Three years ago, he had an ITB pump implanted for spasticity refractory to the high doses of oral baclofen. He had a significant improvement in spasticity, social and functional capacity in the past three years.Later, he developed some disorientation and increased spasticity. He was taken to a local physician who prescribed oral baclofen (120 mg daily in four divided doses) for his increased spasticity. He also advised him to have his ITB pump checked immediately. The following day, his spasticity increased even after taking oral baclofen. He developed multiple seizures and respiratory distress in the next 24-hour period. Subsequently, he was admitted in a local hospital where he was orally intubated and transferred to our ICU for aggressive management.2O; and FiO2, 60%. His spontaneous respiratory rate was 18 breaths per minute and an oxygen saturation of 100% was noted on pulse oximetry. In the local hospital, he was documented to have a high fever of 107°F (41.6°C) and he had received intravenous lorazepam, phenytoin, pantoprazole, piperacillin/tazobactem and dopamine. On physical examination, neurologically he was unconscious with decerebrate posturing and his Glasgow coma scale was 6. He had an absent corneal and gag reflexes. He was moving all four limbs in response to noxious stimuli. He was also noted to have an extreme spasticity in all four limbs. Lung examination revealed decreased breath sounds in the left lower base. Cardiac examination was unremarkable. He had a palpable baclofen pump on abdominal wall and bowel sounds were heard. The differential diagnoses were septic shock, meningitis, neuroleptic malignant syndrome and malignant hyperthermia.On presentation, his temperature was 104.6°F (40.3°C), heart rate 127 beats per minute, and the blood pressure was 85/45 mm/Hg. His ventilator settings were: assist-control ventilation mode; respiratory rate, 15 breaths per minute; tidal volume, 650 mL; positive end expiratory pressure (PEEP), 5 cm HThe initial laboratory results showed serum creatinine phosphokinase (CPK) 5250 U/L  and CPK-MB fraction 12.1 ng/ml . Serum chemistry revealed sodium 142 mmol/L, potassium 5.1 mmol/L, chloride 120 mmol/L, bicarbonate 13 mmol/L, and creatinine 2.1 mg/dl. Hemogram showed white blood cell count 12.2 K/UL, hemoglobin 16.5 g/dl and platelet count 9 K/UL . Liver function test showed aspartate aminotransferase (ALT) 1128 U/L, alanine aminotransferase (AST) 1140 U/L, alkaline phophatase 90 U/L, total bilirubin 1.2 mg/dl, conjugated bilirubin 0.7 mg/dl, prothrombin time 20.2 seconds , and INR 2.0 . Blood and urine cultures were obtained. Chest radiograph was normal. A computed tomography (CT) scan of the chest revealed atelectasis of the left lung base. His CT scan of head did not show any acute infarct or bleeding. His initial management included intravenous fluids, norepinephrine, platelet transfusion, phenytoin, propofol and broad-spectrum antibiotics (vancomycin, ceftriaxone) for suspected meningitis and septic shock. He received intravenous lorazepam (4–8 mg every four hours) for his spasticity. Next day, his spasticity improved and an ITB specialist investigated his baclofen pump. His baclofen pump analysis revealed that it was stopped due to some programming error, which was restarted at a previously prescribed baclofen rate (260 μg/day).On third hospital day, his serum CPK was 15,878 U/L, AST was 2566 U/L, ALT was 2993 U/L, while CPK-MB fraction came down to 3.4 ng/ml. His urine output decreased (<400 ml/ day) and serum creatinine increased in the range of 5–6 mg/dl. Later, he was hemodialyzed few times during the course of hospitalization due to acute renal failure. His echocardiogram showed left ventricular ejection fraction of 20–25% and severe global hypokinesis. His electroencephalogram did not reveal any epileptogenic activity. He developed full-blown multisystem organ failure with an evidence of shock liver, renal failure, respiratory failure, disseminated intravascular coagulation and myocardial depression. His nutrition was started on nasogastric tube feedings, and proper ventilator care was taken through a tracheostomy tube. His serum baclofen obtained at the time of admission was less than 0.02 μg/ml . After a three-week course of aggressive management in ICU, he was weaned off from the ventilator and his multiple organ shock resolved. At a six-month follow-up, he was observed in a nursing home with his baseline functional, social, and family activities.Intrathecal baclofen provides an effective improvement in the spasticity of patients whose spasticity is not sufficiently managed by oral baclofen or other oral anti-spastic medications . RegardlB receptor to cause muscle relaxation. Baclofen reduces increased muscle tone, Babinski sign, tendon reflexes, ankle clonus and sometimes decreases muscle force.The precise mechanism of action of baclofen as a muscle relaxant and anti-spasticity agent is not fully understood. Baclofen inhibits both monosynaptic and polysynaptic reflexes at the spinal cord level , possiblB receptors in the CNS and spinal cord. Down-regulation of GABAB receptors accounts for the decreased sensitivity to the baclofen over time. Although GABAB receptors are down-regulated, it is the baclofen itself that causes increased inhibitory tone in the CNS and spinal cord [Long-term ITB infusion causes down-regulation of GABAnal cord . Therefonal cord -12, rhabnal cord , cardiacnal cord ,12,13 hanal cord . Close aA receptors of spinal cord by different mechanisms [B receptors do not interfere with benzodiazepine's mechanism of action. During a planned removal of ITB pump due to infection or other causes, premedication with high doses of benzodiazepines and augmented oral baclofen is usually administered in the hospitals to prevent spasticity. Similarly, high doses of oral baclofen is also tried in some cases of ITB withdrawal syndrome [B receptors due to chronic ITB administration. Moreover, it has been suggested that it may take many hundreds of grams of oral baclofen to achieve a therapeutic baclofen level in the cerebrospinal fluid, compared to the patients who had effective spasticity control with an ITB pump [B receptor agonist by prompt restoration of ITB pump and proper supportive care might be the best treatment. Similarly, high-dose benzodiazepines may be effective because of similar mechanism of action on widespread CNS GABAA receptors. High-dose benzodiazepines could be an initial life saving strategy even before analysis and restoration of ITB pump is achieved, or in cases, where resumption of ITB administration is not as simple as correcting a programming error.Benzodiazepines are helpful in controlling spasticity and seizures during ITB withdrawal syndrome ,10. Benzsyndrome ,16. But syndrome . High doITB pump . AlthougB receptor-mediated presynaptic inhibition of serotonin [Intrathecal baclofen bolus is appropriate, but due to the risk of inadvertent overdose, an experienced physician should immediately perform reinstitution of ITB pump. We had restarted ITB in our patient at a previously prescribed dose. However, a much higher dose of baclofen could have safely been given to overcome the spasticity since the patient was in an ICU. High-dose dantrolene  has been tried to reduce spasticity and fever in ITB withdrawal syndrome . Reductierotonin . We did Baclofen withdrawal syndrome is a potentially life-threatening complication of intrathecal baclofen pump. Empty pump reservoir, catheter leaks or displacement, pump malfunction, programming error and refill of pump with improper drug concentration are the possible mechanisms which could lead to an ITB withdrawal syndrome. Regular check-up of the ITB pump by a specialist, educating patients and their caregivers may decrease the incidence of ITB withdrawal syndrome. Oral baclofen replacement may not be an effective method to treat or prevent ITB withdrawal syndrome. Early recognition of syndrome, high-dose benzodiazepines, prompt analysis of the ITB pump with reinstitution of baclofen, and proper intensive care management are mainstays for the management of ITB withdrawal syndrome.GABA -gamma amino butyric acid, ITB – intrathecal baclofen, CPK – creatinine phosphokinase, ALT – aspartate aminotransferase, AST – alanine aminotransferase, CT – computed tomography.None declared.IM: Direct patient care, article conception, critical and extensive revision of article for important intellectual content, review and drafting of the original article. AH: Literature search, case review and summary, drafting of the original article. Both authors read and approved the final manuscript and contributed equally to the manuscript.The pre-publication history for this paper can be accessed here:


It has recently been argued that unsafe medical injections are a major transmission route of HIV infection in the generalised epidemics of sub-Saharan Africa.We have analysed the pattern of injections in relation to HIV incidence in a population cohort in Manicaland in a rural area of Zimbabwe. In Poisson regression models, injections were not found to be associated with HIV in males  or females .It is important that unsafe medical injections can be confidently excluded as a major source of HIV infection. In rural Zimbabwe the evidence is that they can. A debate is raging over whether unsafe medical injections are an important cause of HIV infection in Africa; in rural Zimbabwe evidence suggests they are not The widely held belief that heterosexual transmission is the driving force behind sub-Saharan Africa's (SSA's) HIV epidemic  has recePresently, the only published data on the possible contribution of injections to HIV transmission in SSA come from rural Uganda, where Kiwanuka et al. demonstrated no association of injections with HIV incidence . Data frIn this paper, we analyse data from a population cohort in Manicaland in rural Zimbabwe. We describe the determinants of injections in adults and then test the association between injections and incidence of HIV infection.n = 1,606; 83.6%) was used for all analyses except for the examination of rates of injections stratified by HIV status at baseline.Data were analysed from the baseline (1999/2000) and follow-up (2002/2003) rounds of a cohort of adults in the Manicaland HIV/STD Prevention Study. Eligible men and women aged 15 to 54 were recruited based on an initial household survey . In respAt follow-up, participants were asked whether they had received an injection or had been pricked by a needle since the baseline interview. A range of health and socio-demographic data were also collected, including self-reported history of sexually transmitted disease (STD) symptoms. Reports on STD symptoms were from the 1-y period before the follow-up interview, and thus did not correspond to the entire 3-y follow-up period. HIV serological testing was performed on dried blood spots using a highly sensitive and specific antibody dipstick assay .Ethical approval for this study was granted by the Medical Research Council of Zimbabwe (MRCZ/A/681) and the Applied and Qualitative Research Ethics Committee, Oxford University, United Kingdom (N97.039). Written informed consent was sought from study participants.p-value ≤ 0.1. Using the same strategy, models were then fitted with HIV as the outcome variable.The objective of this analysis was to test the plausibility that injections are an important risk factor for HIV incidence. Having received an injection was modelled as a proximate cause with demographic variables, sexual behaviour, and STDs acting as potential confounders. Determinants of injections were analysed with univariable and multivariable Poisson regression models of incidence rate ratios (RRs). Attributes were retained in age-adjusted multivariable models if the stratum-specific RR differed from one and had a Wald-test p = 0.81 for males; RR = 1.13, p = 0.28 for females)  reported having received an injection or needle prick during the 3-y follow-up period. Females reported more injections than males (RR = 1.93). Rates were not significantly higher for individuals who were HIV positive at baseline  .Being widowed, separated, or divorced was the only attribute associated with increased rates of injections for males . For femThere were 67 HIV seroconversions ; 13 (19%) of those seroconverting reported not having had sex in the inter-survey period, and 40 (60%) reported not having had an injection during the period .There was no significant association between injections and HIV incidence among either males or females—in either unadjusted or adjusted models. For males, HIV seroconversion rates were elevated amongst 25- to 44-y-olds, sexually active individuals, and those who had suffered STD symptoms, though none of these attributes reached levels of statistical significance . For femThese data, from a population cohort in rural Zimbabwe, suggest that—at the population level—injections are not a major route of HIV transmission. There was a very slight, non-significant association between injections and HIV amongst females. Could this association achieve significance with a greater sample size or more events? Given the strong association between STD symptoms and both injections and HIV incidence, this is possible, but would likely be a result of residual confounding. In other words, both HIV and injections have a common association with STDs. It has been argued that the association between sexual activity and HIV is confounding between STDs and pregnancy and injections and that it is the injections that are causal. Our analysis does indeed find these associations—but finds STD symptoms the strongest predictor of new HIV infections.Our measure of injection risk is unambiguous but lacks many dimensions relating to unsafe injections. Presently, we have collected data about the receipt of injections and other needle pricks. Thus, the exposures in these analyses are not restricted to injections received from the health-care sector—the source that Gisselquist et al. originally hypothesized as a major route of transmission . Also, tn = 4) where individuals seroconverted who had reported never having had sex may still be a product of social desirability reporting bias. Clearly, incidence data offer the most explanatory power in elucidating the determinants of the HIV epidemic, but these anomalous cases also highlight the difficulties of collecting time-varying sexual behaviour and serostatus information.Thirteen of 67 individuals seroconverting in this study reported no sexual partners in the inter-survey period. Only four of these 13 reported never having had sex. This leads us to suspect that incorrect categorisation of HIV status at baseline in addition to misreporting of sexual behaviour may explain some of these infections: it is possible that a proportion of the nine individuals who reported having had sex in their lifetime but not in the inter-survey period had been recently infected but had not yet seroconverted at the time of the baseline testing . In thisInterestingly, in light of expected HIV-associated disease and care, individuals who were HIV positive at baseline did not have higher rates of injections than individuals who were HIV negative. Injections were found to be highly associated with childbirth/pregnancy. HIV-positive women—and especially those at advanced stages of infection—are known to experience reduced fertility . TherefoHad injections proven to be a risk factor for HIV incidence, further investigations would have been needed to determine the source and types of risky needle pricks. However, no such association was found. Unsafe injections are unacceptable, but this evidence suggests that they do not play a major role in the transmission of HIV in rural Zimbabwe.There is a lot of controversy over whether the spread of HIV in sub-Saharan Africa is predominantly caused by unsafe sexual practices, or whether unsafe medical injections given by health professionals might also have a prominent part to play. A recent paper suggested that unsafe medical injections were important.In an ongoing survey in rural Zimbabwe between 1999/2000 and 2002/3 that was trying to assess why some individuals get infected with HIV, the authors asked 505 men and 1,342 women a number of questions. They asked them about their sexual history, whether they had children, and whether they received injections. They tested the adults for HIV at the beginning and end of the study period.744 people had had a medical injection and 67 people acquired HIV. There was no evidence overall that injections were linked with an increase in HIV infection. The strongest link with HIV infection was with symptoms of sexually transmitted diseases (in other words, people with these symptoms were more likely to acquire HIV infection).These findings suggest that although it is still possible for an individual to get HIV through unsafe medical injections, overall in this population in Zimbabwe, unsafe injections are not an important cause of HIV infection. Hence policymakers should concentrate more on trying to prevent infection from unsafe sex.http://www.thebody.com/safesex.htmlInformation on safe sex: http://www.who.int/hiv/topics/harm/reduction/en/World Health Organization Web page on reducing the risk of HIV infection in drug users who inject drugs intravenously: http://www.unaids.org/html/pub/publications/fact-sheets01/Zimbabwe_en_pdf.htmFact sheet from the Joint United Nations Programme on HIV/AIDS on HIV/AIDS in Zimbabwe:


Rauwolfia serpentina was widely used for its antihypertensive action in the past. In later years, its use has been reduced because of precipitation of depression and extra pyramidal symptoms due to its central action. In the present investigation, reserpine methonitrate (RMN), a novel quaternary analogue of reserpine was synthesised and evaluated biochemically for its central and peripheral amine depleting actions in rats while its influence on the blood pressure was measured in anaesthetized rats in comparison with reserpineReserpine, an alkaloid from Reserpine treatment (5 mg/kg) produced a significant increase in the urinary excretion of VMA, 5-HIAA and HVA while RMN at doses of equal to and double the equimolar doses of reserpine (5 and 10 mg/kg) produced significant increase in VMA and 5-HIAA excretion without producing any effect on HVA excretion compared to control animals. Reserpine in the dose range of 0.5 to15 μg/kg produced significant reduction in blood pressure compared to control. RMN was also found to produce significant decrease in blood pressure at doses of 10, 25 and 50 μg/kg body weight in comparison to control. The results indicated peripheral depletion of biogenic amines by RMN without affecting the central stores of the amines.The present study clearly indicated that the quaternization of reserpine restricts its transfer across the blood-brain barrier and could be the reason for its selective peripheral action. It is also clear that the hypotensive actions of RMN could be due to peripheral depletion of catecholamines. Rauwolfia species, was introduced for the treatment of hypertension and schizophrenia in 1950's but was replaced by more effective drugs by the end of 1970's [Reserpine, an alkaloid isolated from f 1970's . Reserpif 1970's . Howeverf 1970's -17. As af 1970's ,19. HencAttempts were made in the past to synthesize derivatives of reserpine with possibly higher and/or modified activities or with fewer side effects -23. Compviz., reserpine methonitrate (RMN), which was synthesized in our laboratory was evaluated in rats for its amine depleting actions compared to reserpine. For this, the urinary levels of vanillylmandelic acid (VMA), 5-hydroxyindoleacetic acid (5-HIAA) and homovanillic acid (HVA) which are the respective metabolites of noradrenaline, serotonin and dopamine were estimated after reserpine or RMN treatment in rats. The change in the blood pressure response of anaesthetized rats after treatment with RMN was also evaluated in comparison to reserpine.Based on the poor ability of quaternary compounds to penetrate the blood-brain barrier, a great deal of research has been devoted towards quaternization of existing drugs to achieve preferential peripheral action -32. EarlThe main aim of the study was to determine whether RMN was able to deplete the central and peipheral biogenic amines to the same extent as produced by reserpine. Reserpine at a dose of 5 mg/kg body weight produced significant increase in the urinary excretion profile of VMA compared to control animals. The analogue at doses equimolar to reserpine of 5 and 10 mg/kg body weight produced more significant increase in VMA excretion compared to controls and that observed with reserpine  of RMN did not produce any further increase in the VMA and 5-HIAA excretion compared to lower dose. The possible reason for this effect could be that 5 mg/kg dose was sufficient to deplete the amines completely from the storage sites.In order to evaluate whether the quaternary analogue of reserpine (RMN) still retains the peripheral blood pressure lowering activity, further experiments were carried out on the blood pressure of anaesthetized rats. Thus far, the results of RMN on the blood pressure response of anaesthetized rats confirmed that the peripheral actions of reserpine molecule are not affected by quaternization. However, in the present study the vehicle (DMSO) also produced minor hypotensive effect on blood pressure of rats when administered alone with the dose used for the administration of the drugs. Earlier workers  also repThe effect of equimolar doses of RMN also indicated hypotension however, with higher doses compared to reserpine. It is further indicated that quaternization of reserpine not only restricted the entry of RMN to central nervous system but also reduced to the target tissue in the periphery. Hence relatively higher doses were required to produce reserpine like effect. Mechanistically, the hypotensive actions of RMN could also be due to peripheral depletion of catecholamines as evident from the positive correlation with the results of previous section on the peripheral depletion of monoamines.In conclusion, the present study indicated that the quaternization of reserpine molecule prevents its access into the central nervous system and thereby produces selective peripheral depletion of biogenic amines. Furthermore, the study indicated that quaternization of reserpine had not abolished the hypotensive response but only higher doses were required.The synthesis of RMN was done as follows: The solution of reserpine (2 g, 3.3 mmols) in dichloromethane (20 ml) was added to methyl iodide (11 ml, 176 mmols) and the resulting mixture was kept for two days in dark. The solid was filtered and washed with a little cold dichloromethane and dried under vacuum at 70°C for 2 h to yield reserpine methiodide (RMI) ,34. ThenReserpine and thiopentone were generous gift samples from Novartis India Limited and Abbott Laboratories, Mumbai respectively. The standard samples of VMA, 5-HIAA, HVA and iso-VMA  were purchased from Sigma-Aldrich, St. Louis, USA. All other chemicals used were of HPLC or analytical grade as appropriate.The solutions of reserpine and RMN under study were prepared in DMSO and the volume of each dose was adjusted to 0.1 ml/100 gm body weight as suggested by Varma et al., . The dosad libitum with 12 h light and dark cycle. The animal experiments conducted in this research work were approved by the Institutional Animal Ethics Committee and by the government regulatory body for animal research (Regd. No. 516/01/A/CPCSEA).Albino rats of either sex weighing between 100–150 gm (Charkaborty Enterprise, Kolkata) were used in the study. They were acclimatized to the laboratory conditions for at least 10 days prior to the experiment and were provided with standard diet and water Animals were divided into 4 groups of six each and were housed individually in metabolic cages. Funnels of suitable size were arranged at the bottom of the metabolic cages for collection of urine. Perforated plastic discs were arranged in the funnels to retain fecal matter. The animals were maintained at room temperature and acclimatized to metabolic cages for few days prior to drug administration.The treatment given to the groups of animals was as follows:Group 1: Control animals treated with DMSO intraperitoneally at a dose of 0.1 ml/100 gm body weight.Group 2: Animals administered intraperitoneally with reserpine at a dose of 5 mg/kg body weight.Group 3: Animals administered intraperitoneally with RMN at a dose equivalent to 5 mg/kg body weight of reserpine.Group 4: Animals administered intraperitoneally with RMN at a dose equivalent to 10 mg/kg body weight of reserpine.M HCl arranged at the bottom of the funnel. The volumes of the 24 h urine samples collected in the beakers were noted individually and about 2 ml of urine (mixture) from each animal was taken into sample tubes and centrifuged at 3000 rpm for 10 minutes. The supernatants were transferred into another set of clean and dry tubes and stored at -20°C until analysis by HPLC.In each group, animals were placed individually in metabolic cages after drug administration and were allowed access to water. The 24 h urine samples from the point of drug administration was collected for each animal in a beaker containing 5 ml of 6 M KH2PO4 and the flow rate of the mobile phase was maintained at 0.8 ml/min, which yields a column back pressure of 220–230 kgf/cm2. Detection was done by UV absorption at 230 ηm. The range of the detector was set at 0.001 a.u.f.s. The peak area ratios of VMA, 5-HIAA and HVA to that of internal standard were calculated and substituted in the respective regression equations to estimate the amount of the metabolite present in the sample.The procedure described by Wako-chem. Co., was usedThe procedure described by Noble  was follHaemodynamic setup was used to record the blood pressure of rats. The blood pressure of each animal was recorded from left common carotid artery connected to a mercury manometer on kymograph paper. The normal blood pressure of rats was recorded after stabilization for 30 minutes. The different doses of reserpine (0.25, 0.50, 1, 5, 10 and 15 μg/kg body weight) or RMN (10, 25 and 50 μg/kg body weight) were studied in separate groups (n = 6) to determine the change in blood pressure response.t-test. In all the cases, p < 0.05 was considered statistically significant.Data are expressed as mean ± standard error of means. Statistical analysis was done using one-way analysis of variance (ANOVA). Post-hoc comparisons were done by using Dunnet's RMN: Reserpine methonitrateVMA: Vanillylmandelic acid5-HIAA: 5-Hydroxyindoleacetic acidHVA: Homovanillic acidDMSO: Dimethyl sulfoxideSN made significant contribution in designing the studies, conducting the experiments, interpretation of the data, conceptualization of statistical analyses and drafting the final manuscript. KMB assisted in experimental work, data analysis and writing of the manuscript. SS conceived the study, made substantial contributions in data analysis, data interpretation, writing of the manuscript and in coordination of the experiments. All authors read and approved the final manuscript.


Axix axis) as a prey species, dhole or Indian wild-dog  as the predator and a protozoan (Sarcocystis axicuonis) as the parasite. We examine here, with the help of a model, the ecological conditions necessary for the evolution and stability of such a mutualistic relationship. A two – level game theory model was designed in which the payoff of a parasite is decided not only by alternative parasite strategies but also by alternative host strategies and vice versa. Conditions for ESS were examined.For parasites with a predator-prey life cycle, the completion of the life cycle often depends on consumption of parasitized prey by the predator. In the case of such parasite species the predator and the parasite have common interests and therefore a mutualistic relationship is possible. Some evidence of a predator-parasite mutualism was reported from spotted deer or chital (A tolerant predator strategy and a low or moderately virulent parasite strategy which together constitute mutualism are stable only at a high frequency of recycling of parasite and a substantial prey – capture benefit to the predator. Unlike the preliminary expectation, parasite will not evolve towards reduced virulence, but reach an optimum moderate level of virulence.The available data on the behavioral ecology of dhole and chital suggest that they are likely to meet the stability criteria and therefore a predator-parasite mutualism can be stable in this system. The model also points out the gaps in the current data and could help directing further empirical work. Cuon alpinus) and a protozoan parasite (Sarcosystis axicuonis) with chital or spotted deer (Axis axis) as the prey-host [Preferential killing of sick and disabled prey individuals by the predator has been the focus of many ecologists working with different predator – prey systems. In a variety of prey predator systems, diseased or weaker animals are shown to be consumed in greater proportion by predators -5. Increrey-host ,11.Sarcocystis system makes group selection and viscosity very likely factors in shaping the relationship [There can be a potential problem in such a mutualistic relationship. Low virulence of the parasite towards the predator host and parasite tolerance by the predator host are essential factors for the maintenance of a mutualistic relationship. However, it is possible that a virulent parasite can grow faster and invade a mild parasite population. On the other hand a parasite resistant predator can avoid the cost of parasitism but share the benefit of prey capture and therefore invade a tolerant population. Either of the events can destabilize the mutualistic relationship. It is essential therefore to examine the evolutionary stability of the mutualism. In a completely randomized distribution, a mild parasite population can be easily invaded by a virulent one and a tolerant predator can be invaded by a resistant one. Population viscosity, group selection and kin selection can alter the evolution of virulence . The dhotionship . The liftionship ,14. TherWe examine here the effect of parasite recycling on the evolution of a predator-parasite mutualism, using a theoretical model.We consider two alternative strategies, namely mild and virulent, for the parasite Table . The virIf the parasite population consists of the mild type, they enjoy a fitness of 'm' from the tolerant host and 'p*m' from a resistant one. Since they exert a cost y on the host, there is erosion of the host resource. The host resource available to them is therefore (1-y). Similarly for a population of virulent parasites the mean fitness gain is v and the host resource available (1-x). A virulent host invading a mild population will gain a fitness of 'v' such that v > m. In the absence of recycling the host resource available to it would be (1-y). However with a frequency of recycling 'fr', the host resource would be,fr (1-x) + (1-fr)(1-y)Similarly, that for a mild parasite invading a virulent population would be,fr (1-y) + (1-fr)(1-x)If the predator population is tolerant the parasite will be harbored in large numbers and disseminated to the prey population. Since the parasitized prey is more susceptible to predation the predator gets a benefit 'z' of easy catching. A resistant population, on the other hand has a small probability 'p' of harboring the parasite. Therefore the benefit the predator gets would be 'p*z'. If a resistant predator invades a tolerant population, with the recycling factor 'fr', the benefit of prey capture would be,fr * p *z + z(1-fr).The benefit for a tolerant one invading a resistant population would be,fr * z + (1-fr)* p*z.We assume that 'v', the benefit for a virulent parasite by infecting single host is directly proportional to 'x' i.e the loss to the host from infection by virulent parasite.V = α *x similarly, m = α *yA mild parasite will be able to invade a virulent population if the pay-off to the mild invader is greater than that for the virulent population. When the predator population is tolerant, this condition is satisfied when,y * [fr (1-y)+(1-fr) (1-x)] > x (1-x)The condition under which a virulent invader is unable to invade a mild population is(1-y)* y > [fr *(1-x) + (1-fr)*(1-y)]*xSince x > y, for satisfying both these conditions, fr should be large, y should be moderate and x should be large. Thus selection would favour a moderate virulence in the parasite towards the predator host. Unlike our expectation, low virulence is unlikely to be stable. However, a mutualistic relation can remain if the prey capture benefit is sufficiently large. It can be easily seen that the above conditions remain unaffected even if the predator population is resistant.Considering predator strategies, a tolerant predator will be able to invade a resistant population in the presence of a mild predator if,p *z - p*y < (1-fr) p *z + fr*z - yi. e. p*fr * z - p* y < fr *z - yThis condition will be satisfied if fr *z > y since p < 1. A resistant predator will be unable to invade a tolerant population since the necessary condition isz-y < (1-fr)*z + fr*(p*z) - (p * y)fr*z - y < p*fr*z - p*yThis invasion is impossible if fr*z > y. If the parasite is virulent, the necessary condition would be fr*z > x. Since x is assumed to be large, a resistant predator would be stable if the prevalent parasite strategy is virulent.Thus when fr and z are large and y is small mutualism would be stable. When this condition is not satisfied the predator will evolve resistance to the parasite and the parasite will evolve greater virulence.A large recycling frequency (fr) appears to be the only critical factor in the evolution of parasite virulence. However, the parasite is unlikely to evolve towards low virulence. There will be a moderate virulence optimum. For a net benefit to the host the cost associated with this level of virulence should be less than the benefit in terms of ease of prey capture.For the evolution and stability of the tolerant strategy in the predator a large fr as well as large z and small y are necessary. A predator-parasite mutualism therefore critically depends upon these factors, whereas it is independent of p and α.Sarcocystis system, these condition are very likely to be satisfied. Field data show that the frequency of sarcocystosis of the heart in dhole kills was approximately double that of chital dyeing of other causes [2 [S. axicuonis show no apparent symptoms of disease or abnormality [S. axicuonis towards dhole is low. Further, the division of labor can substantially reduce the effective cost of carrying the parasite. Currently we are unable to quantify these parameters empirically and have no estimate of the actual cost of harboring the parasite. Therefore we are unable to state quantitatively that all the necessary conditions for mutualism are met by the system. The importance of the model is that it helps us identify the gaps in the data and thus orient future empirical work.In the case of Dhole-chital – r causes . This suauses [2 . There iauses [2 . The terauses [2 ,16. Thisauses [2 , the freauses [2 , furtherormality ,20. ThisAlthough we are far from having an empirical estimate of y, fr and z, the known ecology of chital and dhole suggest that fr and z could be sufficiently large. This makes the system a likely candidate for the evolution of predator parasite mutualism. Any other predator-prey system that satisfies these conditions is also likely to co-evolve with some parasite species towards a predator-parasite mutualism. Parasites of diverse taxa have evolved predator-prey life cycles and any of them could be possible candidates for a mutualism. Predator-prey-parasite systems that satisfy the following three criteria are the most likely candidates for a stable mutualistic relationship:i) parasitized prey individuals are killed with substantially greater frequency by the predatorii) pathogenicity of the parasite towards the predator host is low or moderateiii) there is a high rate of parasite recycling to the predator host.Sarcocystis system may not be unique and many possible examples of predator-parasite mutualism may be present in nature.We need to look at a number of systems that could satisfy these criteria. The chital-dhole-Both authors have contributed approximately equally to the model development.