Several HIV prevention programs use data on condom sales and survey-based data on condom prevalence to monitor progress. However, such indicators are not always consistent. This paper aims to explain these inconsistencies and to assess whether the number of sex acts and the number of condoms used can be estimated from survey data. This would be useful for program managers, as it would enable estimation of the number of condoms needed for different target groups.We use data from six Demographic and Health Surveys to estimate the total annual number of sex acts and number of condoms used. Estimates of the number of sex acts are based on self-reported coital frequency, the proportion reporting intercourse the previous day, and survival methods. Estimates of the number of condoms used are based on self-reported frequency of use, the proportion reporting condom use the previous day and in last intercourse. The estimated number of condoms used is then compared with reported data on condom sales and distribution.Analysis of data on the annual number of condoms sold and distributed to the trade reveals very erratic patterns, which reflect stock-ups at various levels in the distribution chain. Consequently, condom sales data are a very poor indicator of the level of condom use. Estimates of both the number of sexual acts and the number of condoms used vary enormously based on the estimation method used. For several surveys, the highest estimate of the annual number of condoms used is tenfold that of the lowest estimate.Condom sales to the trade are a poor indicator of levels of condom use, and are therefore insufficient to monitor HIV prevention programs. While survey data on condom prevalence allow more detailed monitoring, converting such data to an estimated number of sex acts and condoms used is not straightforward. The estimation methods yield widely different results, and it is impossible to determine which method is most accurate. Until the reliability of these various estimation methods can be established, estimating the annual number of condoms used from survey data will not be feasible. Collecting survey data on the number of sex acts and the number of condoms used in a fixed time period may enable the calculation of more reliable estimates of the number of sex acts and condoms used. Programs that promote condom use for HIV prevention typically monitor their progress through survey-based indicators, such as the percentage of the population who ever used a condom or the percentage who used a condom in their last sex act with a casual or regular partner ,2. Such The purpose of this study is to explain inconsistencies between information on reported levels of condom use and data on the number of condoms sold and distributed. Understanding the apparent inconsistencies between sales and survey data will help clarify to what extent the concerns about condom wastage, misreporting, and other related problems are founded. It will also provide guidance for improving the monitoring of condom sales and distribution, and for improving survey questionnaires. To achieve these objectives, we use survey data from six Demographic and Health Surveys to estimate the total annual number of sex acts in a country, and the total number of condoms used in those sex acts, and compare the totals with reported data on condom sales and distribution.Protector Plus condoms increased from 1.9 million in 1997, to 4.8 million in 1998, to 8.9 million in 1999. Data on public sector condom distribution, which we discuss later in this paper, suggest that public sector sales also increased substantially. Yet, nationally representative surveys indicate that condom use in last sex stayed constant between 1996 and 1999 at roughly 34% for males and 17% for females [Salama condoms increased steadily between 1995 and 2000, as did condom distribution by the Ministry of Health. However, survey data indicate that condom use at last intercourse remained roughly constant between 1996 and 1999, for both men and women [At least in some instances, survey information on condom use and condom sales records appear to be inconsistent ,6. For eInaccuracies in the number of condoms sold or distributed are likely because sales figures typically represent sales to the trade  rather than sales to consumers. Consequently, the recorded sales numbers will include condoms that are being stocked at various levels of the distribution chain. In addition, some of the condoms that are sold and/or distributed may be wasted or smuggled to other countries.In addition to these potential problems with condom sales data, there are concerns that reported condom use in surveys may be inaccurate. For example, there are concerns that respondents may overreport condom use because they do not want to admit to the interviewer that they are engaging in risky sexual behavior. There are also concerns that condom use may be underreported because condoms are frequently used with sex workers, which stigmatizes condom use. Women may also underreport condom use because it is a male method. Some questionnaires try to overcome this by asking "The last time you had intercourse, was a condom used?" rather than "The last time you had intercourse, did you use a condom?" .This study uses two types of data: data on condom sales and distribution, and survey data on self-reported condom use. We restrict our analysis to data from four countries in sub-Saharan Africa , largely because these countries have strong condom social marketing programs and therefore relatively good data on condom sales and distribution. In addition, Tanzania and Zimbabwe are two of the countries where discrepancies between condom distribution and condom use have been noted.Data on sales of socially marketed condoms were obtained from DKT International's Social Marketing Statistics -14, whilThe survey data used in this study include the following Demographic and Health Surveys (DHS): Kenya 1998), Nigeria (1999), Tanzania , and Zimbabwe . Each of the six surveys comprises a representative sample of females aged 15–49 and of males 15–54 , such a question has not been included in recent surveys . For exaAll DHS surveys asked whether respondents used a condom in their last sex act. We use this information to estimate the probability of condom use, and, subsequently, to estimate the total annual number of condoms used in the country.C) equals the product of the frequency of intercourse, or the number of sex acts (F), and of the probability of condom use (p):In theory, estimating the total number of condoms used in a population is straightforward. The estimated mean number of condoms used per sexually active person (C = F × p     (1)CT) then can be calculated by multiplying C with the proportion of individuals who are sexually active (s) in the population at risk and with size of the population at risk (N):The total number of condoms used (CT = N × s × C     (2)Since the frequency of intercourse and the probability of condom use are known to vary by age and marital status ,27,29-32wa is the weight for age group a, ma and (1 - ma) are the proportion of married and unmarried respondents in age category a, and sam and sau are the proportion of sexually actives where the subscripts am and au refer to the rates for married and unmarried respondents in age category a, respectively.where We used five-year age categories, and based the age weights on the age distribution within the household file of the DHS, as no other reliable data on the age structure of the population in these countries were available . Marital status and the marital status weights were derived from the individual respondent files of the DHS. Following the DHS definition, we define marriage as formal marriage or living together. Information on current sexual activity, defined as having had sex at least once in the past year, was also obtained from the individual respondent files of the DHS. Data on the countries' population size were obtained from the 2003 World Bank World Development Indicators and are summarized in Appendix see .F and p are not readily available and need to be estimated. The following sections describe the procedures for estimating them.Although the above procedure is simple, data on the two main components, This section describes methods to estimate frequency of intercourse. Three types of estimation methods are presented: 1) estimation based on the reported frequency of intercourse during a four-week period, 2) methods based on the proportion of respondents reporting intercourse the day before the interview, and 3) survival analyses based on the time since last intercourse.All methods follow a similar strategy: 1) Estimate the mean likelihood or frequency of intercourse for a specific time unit  for each of the subpopulations, and 2) estimate the mean frequency of intercourse per year for the entire population by calculating a weighted average of the subpopulation results. The general formula is:Fi stands for the annual frequency of intercourse estimated by method i, fiam and fiau for the estimated mean likelihood or frequency of intercourse per time unit using method i for married and unmarried persons in age category a, respectively, and ni the number of time units for this method in a year.where Some surveys asked married respondents separate questions about the time since last intercourse with the respondents' spouse and with the respondents' other partners. Such questions were included in the 1998 Kenya and 1996 Tanzania DHS surveys. For these surveys, the formula becomes:b subscript in Fib indicates that for married respondents marital and extramarital sex were included separately.where the n1 = 13). However, because few recent surveys contain this type of information, it is generally necessary to use other estimation methods.When self-reported data on the frequency of intercourse during the past four weeks are available, such as in the 1994 Zimbabwe DHS survey, the annual number of sex acts can be estimated by extrapolation. Assuming the past four weeks are representative of the respondents' behavior, the mean annual number of sex acts can be estimated by multiplying this four-week frequency with 13 , so that the frequency of intercourse may be slightly underestimated. In turn, the impact of this more frequent intercourse on condom use may be somewhat greater than results would indicate, as the uncounted numbers may represent commercial sex workers with a relatively high condom use.F2 estimates for these surveys appear to be unrealistically low.Another problem with this method is that for some surveys the percentage of respondents reporting last having intercourse the day before the survey does not appear to be reliable. For example, in the 1998 Kenya survey the percentage of respondents reporting last having sex one day before the survey was smaller than the percentage last having sex two days before the survey (4.1% vs. 8.9%). Similarly, in the 1999 Nigeria survey 0.7% reported last having intercourse one day before the survey, compared to 10.0% who reported having sex two days before the survey. In the other surveys, the percentage reporting last having sex the day before the survey is slightly higher than the percentage last having sex two days before the survey. While it is unclear why so few respondents in the Kenya and Nigeria surveys reported last having intercourse the day before the survey, the implication is that the A third alternative is to estimate frequency of intercourse based on data on the duration since last intercourse, which is collected in all DHS surveys ,28. ThisSlaymaker and Zaba  deal witOne of the main weaknesses of this approach is the assumption that daily probability of intercourse is constant and can be estimated with an exponential decay function. Since data on the actual distribution of the intervals between two successive sex acts are not available in DHS surveys, one cannot determine whether the exponential decay function provides a good fit for the data. Using a function that does not match the data well would introduce a very large error in the estimated annual number sex acts (and consequently in the estimated number of condoms used), rendering the results meaningless.As most DHS surveys only contain data on whether a condom was used in the respondent's last intercourse, we must assume that condom use at last sex is typical for the likelihood of condom use for a given subpopulation. Three different estimations for the likelihood of condom use are explored in this paper, two of which are based on data on condom use at last intercourse and one of which is based on the self-reported frequency of condom use.For surveys that collected information on the frequency of condom use, this information can also be used to estimate the probability of condom use. Unfortunately, none of the DHS surveys asked direct questions about both the number of sex acts and the number of condoms used  who reported using a condom at last intercourse. This estimate was also used by Collumbien et al. . InformaAn alternative measure of the probability of condom use equals the proportion of respondents who reported using a condom at last sex among those who had sex the previous day. This indicator has the advantage that it is less likely to be subject to recall errors. It also avoids the problem that condom use at last intercourse may be dependent on the time since last intercourse. However, this measure has the disadvantage that it tends to be less reliable because it is based on information from a much smaller number of observations (those reporting intercourse the day before the interview).We estimate the annual number of condoms used by multiplying the annual number of sex acts with the probability of condom use for each of the strata by age and marital status, as described in Equation 3. Because we have three different methods to estimate the annual number of sex acts and three methods to estimate the probability of condom use, up to nine estimates of the annual number of condoms used are provided, depending on the available data. Moreover, separate estimates were calculated using data from the female and male DHS surveys, as there are known gender differences in the reported frequency of intercourse and levels of condom use ,32,34.Figure Figure Trust condoms increased from 25% in 1998 to 32% in 1999. Similarly, the percentage of retail outlets selling public sector condoms increased from 2% to 6%. The percentage of retail outlets selling other brands stayed constant at 3% [It is clear that these drastic fluctuations in the number of condoms sold or distributed do not reflect real differences in the level of condom use, as this would require major changes in behavior (and behavior is known to change very slowly). Since statistics on the number of condoms sold or distributed reflect sales to the trade, not consumers, it is highly likely that the observed fluctuations in the number of condoms distributed simply reflect fluctuations in condom inventory due to a stock-up of condoms at one or more levels of the distribution system, the addition of new condom outlets, and so on. For example, data from condom distribution surveys in Kenya indicate that the percentage of retail outlets that were selling socially marketed nt at 3% ,36. AssuIn addition, our estimates of the number of public sector condoms are not the actual number of public sector condoms distributed to the population, but rather the total number of condoms provided to each country by international donors. It is possible that many of these condoms are still stocked at Ministry of Health warehouses and similar distribution hubs, or at local health clinics. The actual number of public sector condoms that reach the hands of consumers is unknown. Therefore, the data that are available on the number of condoms that have been sold or distributed seem to provide an estimate of the total of number of condoms that were in circulation during the course of the year, rather than the number provided to consumers.In other words, the current data on the number of condoms sold or distributed provide a very poor estimate of the actual number of condoms used. For example, as shown in Figure F1, with the two estimates based on the duration since last intercourse (F2 and F3). Next, we discuss the results for the other surveys, for which only methods F2 and F3 could be estimated.Table F1) give the highest estimates. Using this method, it is estimated that in 1994, sexually active unmarried males in Zimbabwe had 21 sex acts per year, while sexually active married men had 82 sex acts per year. For females, the number of sex acts is estimated at 9 per year for unmarried females and 82 for married females. This latter finding is fairly consistent with Brown (2002), who estimated the coital frequency for sexually active married women at 7.9 acts per month, which translates into 95 acts per year.The results from the 1994 Zimbabwe survey show that the three estimation methods yield very different estimates of the annual number of sex acts. Estimates based on the self-reported number of sex acts in the past four weeks (F2), which is based on the proportion of respondents who reported having intercourse the day before the interview, results in an estimate of 8 sex acts per year for unmarried males, 61 for married males, 9 for unmarried females, and 59 for married females. Thus, this estimate consistently yields a lower estimate of the number of sex acts than the estimate based on the self-reported frequency of intercourse. This difference appears to be especially large for unmarried males.The second estimation method (F3), which is based on a survival analysis using the assumption of a constant hazard, yields substantially lower estimates of the per capita annual number of sex acts. For unmarried males, the annual number of sex acts is estimated at only 4, while for married males it is estimated at 17. For females, the corresponding numbers are 3 and 10 per year, respectively. These estimates do not appear to be realistic.The third estimation method (F2) and those based on the survival analysis with the assumption of a constant hazard (F3). The results confirm that this latter method consistently yields very low estimates of the number of sex acts. For example, among sexually active married males, the estimate of the annual number of sex acts ranges from 7.2 coital acts per year in the 1999 Nigeria survey to 23.3 in the 1999 Zimbabwe survey. For sexually active married females, the range is from 4.6 to 13.8, again in those same surveys. In other words, the results from the survival analysis using the assumption of a constant hazard suggest that in several countries, even married couples have intercourse less than once per month. Method F2 tends to yield higher estimates of the annual number of sex acts, but for both the 1998 Kenya and 1999 Nigeria surveys these estimates are also unrealistically low. In these latter cases, the low estimates are due to the fact that the number of respondents reporting last having intercourse the day before the survey is considerably lower than the number reporting last having intercourse two days ago.For all other surveys examined here, we can also compare the estimates based on the proportion reporting intercourse the day before the survey The results based on the survival analyses appear unrealistic and are inconsistent with the published literature on the frequency of intercourse. For example, a study on coitus in sub-Saharan Africa estimates that the monthly coital frequency among sexually active married women ranges from 3.0 in Ghana to 8.1 for Rwanda , which cIt is important to note that the results of the survival analyses are greatly affected by the type of decay function selected. Preliminary analysis using a Weibull decay function yielded estimates of the annual number of sex acts that are roughly one and a half to two times as high as estimates based on the exponential decay function proposed by Slaymaker and Zaba . Unfortup1 and p2 are nearly identical .The estimates of the probability of condom use are shown in Table p3) are lower than those based on the data from the last sex act (p2). For example, in the 1999 Tanzania survey, the proportion who used a condom in last intercourse is 15.7% for males and 7.3% for females. By contrast, of those who had sex the day before the interview, the proportion who used a condom is only 9.5% and 4.3%, respectively. In part, these low estimates of p3 appear to stem from the fact that only a small number of survey respondents reported having intercourse the day before the interview. Consequently, there are some age groups where none of the respondents reported using a condom (not shown), which substantially lowers the estimate of the overall probability of condom use.When we compare the different methods to estimate the likelihood of condom use we notice that in the overwhelming number of cases the estimates based on the proportion reporting condom use at last intercourse of those who reported sex on the day before the interview  appears unreliable due to the small number of cases. In addition, the survival analyses yielded unrealistically low estimates of the frequency of intercourse (F3) that appeared inconsistent with the literature. Therefore, estimates that are based on these two factors are unlikely to be reliable. Table F3 and p3 usually yield the lowest estimates of the total number of condoms used.It is unknown which of the estimates is most accurate. However, as we previously noted, the F2 and p2 are likely to be the most reliable. Data from the 1994 Zimbabwe survey confirm that the estimates based on the self-reported frequency of intercourse (p1) and the percentage who used a condom in last intercourse (p2) yield fairly similar results. This was anticipated, given that self-reported frequency of intercourse was coded as a categorical variable and subsequently quantified on the basis of the percentage who reported using a condom in last intercourse. Table F1 and F2 are also fairly close.When self-reported data are not available, estimates based on Nevertheless, all survey-based estimates of the annual number of condoms used are substantially lower than the reported number of condoms sold for almost every country. The only exception is Kenya, where the high estimate of the total number of condoms used based on the 1998 Kenya DHS is fairly close to the number distributed (10.7 million vs. 11.8 million). For the other surveys, the reported number of condoms sold or distributed tends to be 2.5 to 3.0 times higher than even the highest survey-based estimate of the number of condoms used. Comparison with sales data from the previous year does not resolve these differences.The purpose of this paper was to estimate the annual number of sex acts and condoms used based on survey data, and to compare the latter with data on the annual number of condoms sold and distributed. The ability to estimate the number of sex acts from survey data would be a valuable tool for program managers, as it would enable them to estimate the number of condoms needed. Since the available data on condom sales and distribution measure the number of condoms supplied to the trade rather than to the consumer, survey estimates of the total number of condoms used could also help clarify to what extent data on the number of condoms supplied to the trade reflects actual consumer sales.Analysis of the annual reported number of condoms sold and distributed reveals very erratic patterns. The large year-to-year differences in the total number of condoms distributed clearly do not reflect differences in the number of condoms sold to consumers, nor in the level of condom use, as this would imply major changes in behavior. The latter is unlikely to have occurred, since behavior is known to change very slowly. In other words, the large fluctuations in the number of condoms provided to the trade are likely to reflect fluctuations in condom inventory at various levels in the distribution chain. Because of this, the current data on the number of condoms sold and distributed say very little, if anything, about the number of condoms sold to consumers or about actual levels of condom use.To estimate the annual number of condoms used from survey data, survey questionnaires would ideally ask respondents how often they had sex during a given reference period and how often they used a condom during that period. Considering that using very long reference periods  is likely to cause recall errors, a shorter reference period is preferable. Of the DHS studies used in this paper, only one  asked respondents about the frequency of intercourse during the four weeks preceding the survey. For the other surveys, the frequency of intercourse had to be estimated indirectly on the basis of the duration since last intercourse. Although older data on frequency of intercourse are available for some countries, such data may not provide reliable estimates of current behavior, as the HIV/AIDS crisis and other factors may have influenced sexual behavior.If future surveys are to estimate the annual number of condoms used, then questions enquiring about the total number of sex acts and the total number of sex acts in a fixed time period should be added. For example, recent surveys in Zambia asked about the number of sex acts and the number of condoms used in the past week, which can easily be extrapolated to a one-year period . Asking The results of our survey analyses, which are based on DHS data currently available, show that the estimates of both the number of sexual acts and the number of condoms used vary enormously based on the estimation method used. For several surveys, the highest estimate of the annual number of condoms used is tenfold that of the lowest estimate. While some estimation methods can be disregarded because they yield results that are clearly not plausible, it is impossible to determine which of the remaining methods yield the most accurate results. Until the reliability of these various estimation methods can be established, estimating the annual number of condoms used from survey data will not be feasible.To be able to verify the reliability of the estimates of the number of condoms used, it is necessary to have accurate data on the number of condoms sold and distributed to consumers. In developing countries, such is not feasible, in part due to the lack of standardized record-keeping, and because many condoms are distributed through informal retailers, such as street venders and hawkers, who are unlikely to keep records. For the purpose of testing the feasibility of the estimation methods, it may therefore be more productive to use data from developed countries where retail-level condom sales data are available (assuming such data are not proprietary). Alternatively, it may be possible to test the reliability of the estimates in developing countries, by obtaining the relevant sales data on a smaller scale . However, sales data have the drawback that they do not provide information about the characteristics of the consumers. Consequently, sales data are unable to provide detailed information about program impact.The author(s) declare that they have no competing interests.DM conceived of the study and drafted the manuscript. RVR developed the study design and carried out the statistical analysis. Both authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:This file contains the background data for the calculationsClick here for file


In Leber's congenital amaurosis (LCA), affected individuals are blind, or nearly so, from birth. This early onset suggests abnormal development of the neural retina. Mutations in genes that affect the development and/or function of photoreceptor cells have been found to be responsible in some families. These examples include mutations in the photoreceptor transcription factor, Crx.A Crx mutant strain of mice was created to serve as a model for LCA and to provide more insight into Crx's function. In this study, an ultrastructural analysis of the developing retina in Crx mutant mice was performed. Outer segment morphogenesis was found to be blocked at the elongation stage, leading to a failure in production of the phototransduction apparatus. Further, Crx-/- photoreceptors demonstrated severely abnormal synaptic endings in the outer plexiform layer.This is the first report of a synaptogenesis defect in an animal model for LCA. These data confirm the essential role this gene plays in multiple aspects of photoreceptor development and extend our understanding of the basic pathology of LCA. Photoreceptor cells play a primary role in vision by capturing light energy and converting it into neural stimuli. These sensory neurons are a shared element in all organisms capable of sensing light. In humans, genetic diseases of the eye are common and the primary site of disease is most frequently photoreceptors (for review see -3).Photoreceptors have been well studied, particularly with respect to the biochemistry and physiology of phototransduction. Insight into the development of vertebrate photoreceptors, however, has lagged behind our understanding of function. Only recently have the first molecular mechanisms regulating photoreceptor development been identified . Crx LCA is the most severe genetic disease of photoreceptors  and outer segments . To expaIn this study, an ultrastructural analysis of Crx-/- photoreceptors was carried out. As Crx mutations have been associated with Leber's congenital amaurosis, the findings in this study broaden our understanding of the pathology of this disease. Two prominent pathologies were characterized in the Crx-/- retina: (1) An absolute block in outer segment morphogenesis was noted, with the block occuring at the elongation stage of outer segment formation; (2) Crx-/- photoreceptors exhibited a severe perturbation in synapse formation. This represents the first report of a synaptogenesis defect in an animal model of LCA.Mutations in Crx represent one of a collection of gene mutations that lead to an outer segment formation defect. Homozygous null mutations in the peripherin/RDS gene  or in rhIn developing photoreceptors, an extraordinary growth process occurs whereby the outer segment is generated from the nascent connecting cilium  is expressed in photoreceptors R1–R6. In ninaE null mutants, the rhabdomere, a structure analogous to vertebrate outer segments, fails to develop in R1–R6 photoreceptors [ninaE null flies, a ninaE transgene under the control of a heat shock promoter was subjected to various temperature shifts during development. Heat shock during the normal time of rhodopsin onset resulted in substantial and long-lasting rescue of photoreceptor structure and transient rescue of photoreceptor physiology. However, expression shortly before or after this critical period failed to rescue, suggesting that rhodopsin expression during a discrete window of time in development is essential for proper rhabdomere morphogenesis. This result is consistent with observations in the rat wherein rhodopsin onset occurs with strict timing in the developmental history of most rods in vivo [Drosophila otd, otduvi, photoreceptor morphogenesis is also disrupted [In eceptors , reminiseceptors . An intreceptors  demonstr in vivo . Thus, tisrupted .We found that there are many other genes that are dependent upon Crx. Those that are expressed at a lower level in the Crx-/- retina, such as rhodopsin and peripherin, comprise many that are either enriched or specific to photoreceptors in their expression . They inin vitro models [Drosophila homologue of CRB1, Crumbs, has been implicated in photoreceptor morphogenesis [Finally, polarization of photoreceptors was found to be largely intact, as was ciliogenesis. Another LCA gene, CRB1, and a related gene CRB3, have been implicated in ciliogenesis in o models . The Droogenesis . Howeverogenesis , suggestThe Crx-/- mouse demonstrates the most severe abnormality of photoreceptor synapses reported to date. The peripherin-/- mouse develops a normal complement of photoreceptor terminals which then degenerate as the photoreceptors are lost . Also, sAbnormal photoreceptor terminal formation was noted in a study that examined retinal development in the laminin beta2 chain-deficient mouse . SeveralCrx has been implicated in three photoreceptor diseases that result in human blindness, cone-rod dystrophy2, Leber's congenital amaurosis, and retinitis pigmentosa . The coLCA is a disease in which there is little or no photoreceptor function in infancy; thereby, likely developmental in etiology  for 3 to 4 hours at 4°C. The sclera was then partially removed and the retinas were sliced into small pieces and fixed  overnight at 4°C. These procedures were found optimal for maintaining the structural integrity of the photoreceptor outer segments.After fixing, the tissue was washed 2X in PBS for thirty minutes per wash. The tissue was then postfixed in a 1% osmium tetroxide/1.5% potassium ferrocyanide mixture for 2 hours at 4°C. Staining was carried out for 30 minutes in 1% uranyl acetate in maleate buffer (pH = 6.0) at room temperature followed by 1% tannic acid in 0.1 M cacodylate buffer (pH = 7.4) for thirty minutes. The specimens were then dehydrated and embedded in Epon/Araldite. Thin sections were stained with uranyl acetate and lead citrate, and examined in a Jeol JEM-1200EX electron microscope.Specimens used for SEM required removal of the retinal pigment epithelium (PE), enabling visualization of the outer surface of the neural retina. Retinae from Crx-/-, rhodopsin-/-, RDS, or wildtype eyes were dissected free from PE in a dispase solution and fixed in 1.25% glutaraldehyde and 1.0% formaldehyde overnight at 4°C. Tissue was then washed 5X in cacodylate buffer and dehydrated in ascending grades of ethanol. Tissues were subsequently critical point dried in carbon dioxide. All specimens were mounted and coated with sublimated gold-palladium by the sputtering technique. Micrographs were obtained with a Jeol JSM-35CF scanning electron microscope.EM and ER conducted transmission electron microscopy. EM performed scanning electron microscopy. TF and EM generated, characterized and maintained the Crx-/- mouse line. CC participated in the design and coordination of the study and all data analysis. EM and CC drafted the manuscript. All authors read and approved the final manuscript.


The macronutrient composition of the diets were covertly modified with drinks containing 2.1 MJ of predominantly carbohydrate (Hi-CHO), protein (Hi-PRO), or fat (Hi-FAT). Total energy expenditure was measured for seven days on two separate occasions .Ghrelin, a peptide secreted by endocrine cells in the gastrointestinal tract, is a hormone purported to have a significant effect on food intake and energy balance in humans. The influence of factors related to energy balance on ghrelin, such as daily energy expenditure, energy intake, and macronutrient intake, have not been reported. Secondly, the effect of ghrelin on food intake has not been quantified under free-living conditions over a prolonged period of time. To investigate these effects, 12 men were provided with an Preprandial ghrelin concentrations were not affected by macronutrient intake, energy expenditure or energy intake . In turn, daily energy intake was significantly influenced by energy expenditure, but not ghrelin.Preprandial ghrelin does not appear to be influenced by macronutrient composition, energy intake, or energy expenditure. Similarly, ghrelin does not appear to affect acute or chronic energy intake under free-living conditions. Ghrelin, a peptide secreted by endocrine cells in the gastrointestinal tract, is thought to play a significant role in the regulation of energy balance due to its effects on the stimulation of food intake ,2 and weDespite the evidence indicating a role in acute food intake, little is known about the factors regulating ghrelin and its effects on long-term energy balance in humans. One hypothesis is that ghrelin secretion is up-regulated in periods of negative energy balance and down-regulated in periods of positive energy balance . Since eThe purpose of the present study was to determine how changes in macronutrient composition, energy intake, and energy expenditure affect preprandial ghrelin concentrations, and ghrelin's subsequent effects on food intake.Ghrelin was negatively related to body fat percentage  and BMI , but not body weight . There were no significant body weight changes during the seven day observation periods (2).The composition of the treatment beverages and their contribution to daily food intake is listed in Table Average 24 hour energy expenditure  was 13.9 ± 1.9 MJ/d compared to 12.6 ± 1.6 MJ/d for total energy expenditure , which is an average over-reporting of energy expenditure of 11%. Thus, our assumption that subjects would likely misreport energy expenditure and the values would require adjustment was valid.-1 for Hi-CHO, 2869.5 ± 817.3 pg·mL-1 for Hi-PRO, and 2688.2 ± 755.5 pg·mL-1 for Hi-FAT  and total (including treatment beverages) energy intake (breakfast or entire day), or the interactions between these variables (previous or same day). As a further test, we included energy intake for seven days prior to- and two days after each ghrelin value. None of these days were significant . Individual day and mean 24EE up to the prior 4 four days before each ghrelin measurement was also not significant .Selected and total energy intake for the entire day were significantly influenced by treatment period (P < 0.02), Monday/Friday effect (P < 0.003), Sunday effect (P < 0.03), and 24EE (P < 0.008) Table . Classif2) of the variation in energy intake, still a relatively small percentage of explained variation for a hormone purported to exert a large influence on intake.The partial correlation between breakfast energy intake and ghrelin was 0.07. At 80% power, we could have detected a ghrelin effect if the true partial correlation was a small as 0.36. For powers of 90% and 95%, the true partial correlations would have had to be 0.40 and 0.43, respectively. The partial correlation between total energy intake and ghrelin was even lower than that with breakfast energy intake (r = 0.003). Note that, for a partial correlation of 0.40, ghrelin would have only been explaining about 16% , none appear to play a role in preprandial ghrelin regulation. Similarly, ghrelin did not significantly predict macronutrient or energy intake, despite a power analysis indicating that we would have detected even a moderate effect of ghrelin on intake.Most of the evidence linking food intake and ghrelin comes from single meal, short-term studies. The ingestion of amino acids or a protein meal results in a post-prandial increase in ghrelin -14, wherWren et al. were theAlthough body weight typically increases by ≈ 4.5 kg in men and ≈ 7.3 kg in women over the course of 30 years , the humIn conclusion, it appears that macronutrient and energy intake, and energy expenditure have no effect on preprandial ghrelin. None of the variables measured in this study explain the high daily variability in preprandial ghrelin observed over the course of two-seven day periods. In turn, this study fails to detect the energy intake-stimulating effect of ghrelin, despite carefully measured food intake that lasted more than a week and a study powered to detect even a moderate effect of ghrelin.Twelve healthy, non-smoking men were recruited from the Beltsville, MD area to participate in this study Table . All subad libitum from the menus, and could consume any part or all of a food item, then return the remaining portion to be weighed. BHNRC staff that came into contact with the subjects provided no guidance as to the quantities and/or types of food items chosen. During weekdays, subjects reported to the BHNRC in the morning to eat breakfast, pack selected food items for lunch, then return again in the evening for dinner. Any food taken from the HSF that was subsequently not eaten , was returned the next day, and weighed and recorded. On Friday evenings, subjects were provided with coolers packed with a large amount of food for weekend meals. The weekend coolers provided a wide variety of foods in excess quantities, and subjects were allowed to request additional food items be included. Weekend food could be consumed on either day as long as the subjects logged which day each food item was eaten. All uneaten weekend food was returned on Monday, and weighed and recorded. Although subjects were instructed to consume only food items provided by HSF, they were allowed free access to beverages including caloric, noncaloric and alcoholic beverages. Detailed records of the amount, composition and name brand of beverages was submitted daily. In addition to beverages provided on the menu (milk and juice), both regular and decaffeinated coffee and tea were available at meals.Voluntary food intake was studied continuously for 16 weeks, whereby subjects consumed only foods provided by the Human Studies Facility (HSF) at the Beltsville Human Nutrition Research Center (BHNRC). Subjects choose foods Food items offered in the morning (breakfast and lunch) were presented in a cafeteria-style setting as three different rotating menus, each lasting seven days Table . Some foThe goals of the menu design were to allow detection of macronutrient selection by offering a wide range of carbohydrate, fat- and/or protein-rich foods, and to provide a variety of commonly available foods typical of what many Americans eat. In a research setting it is impossible to duplicate the degree of food choice available in real life. However, more than 300 food items were used to develop menus for this study, and specific requests for food items were incorporated into the menus whenever possible.After each subject selected his desired foods, he presented them to a staff member that recorded the identity and weight of each food item by hand and on a computer (combination of bar code recognition of the food item and hand-entering of the weight). Upon termination of feeding, each subject presented his tray to a staff member that weighed any uneaten food. The accuracy of the food item recording process was verified by comparing the information on the computer with the hand-entered logs. This verification procedure was followed daily, and repeated at the end of the study with all food records. Energy and macronutrient composition were determined by consultation with the USDA Nutrient Database for Standard Reference .ad libitum intake, subjects were randomly assigned to two of three treatments. Each treatment lasted 8 weeks with no break between the periods. The treatments consisted of a daily beverage that contained ≈ 2 MJ/day of predominantly carbohydrate (Hi-CHO), fat (Hi-FAT), or a combination of protein and carbohydrate (Hi-PRO) . The intra- and interassay coefficients of variation (CV) were 5.6% and 7.3%, respectively.Before breakfast and after voiding, body weight was determined weekly on an electronic balance to the nearest 0.01 kg. Body composition was measured by Dual-energy X-ray Absorptiometry .To "capture" daily variations in energy expenditure, we combined a self-reported activity log  and doubTwenty-four hour energy expenditure (24EE) was estimated using a daily recording log method, modified from Bouchard et al. . Briefly2 18O (0.16 g/kg body weight) and 2H2O (0.30 g/kg body weight). Urine samples were collected immediately before the dose and on every morning (second void) for the last seven days of the treatment period. The first sample was collected approximately 24 hr after the dose. Enrichments of 2H and 18O in urine samples were measured by infrared spectroscopy and isotope ratio mass spectrometry, respectively. TEE was calculated using the equations of Weir [Total energy expenditure (TEE) was concurrently measured by the doubly labeled water method as described by Speakman , which p of Weir .Individual daily 24EE values were corrected using the ratio adjustment (notation denoting subjects is suppressed),dayx, corrected = 24EEdayx × (TEE/24EEday 1–7), where24EEdayx is the uncorrected daily energy expenditure value from the activity log for one of the seven days (day X),24EEdayx is one), andTEE is the daily mean energy expenditure estimate using doubly labeled water. Represents a single value during the seven days of measurement , where b0 and b1 are the estimated coefficients of the line fit by regressing the SDs on the means, and yi represents the energy intake data. The other variables were transformed using this same family of transformations. This procedure resulted in homogeneous variances for all variables once transformed, satisfying ANOVA assumptions. We present the data on the original scale in tables and figures for ease of interpretation (unless indicated otherwise).To check the assumption of homogeneous variances necessary for valid F-tests and correct P-values, we used the standard technique of plotting the standard deviations (SD's) against the means for selected energy intake, grouping observations by subject and treatment period. The results of this scatter plot revealed a strong positive linear relationship . The relationships between the SD and mean for macronutrient and energy intake , and 24EE were also positive and significant. This indicated that the SD's (variances) were a function of the mean and that the data needed to be transformed. We followed methods described by Draper and Smith , and useDue to the free-living nature of the subjects, there were three observations (of 168) where (for unknown reasons) a subject's food intake differed greatly from habitual intake due to a skipped meal or meals with low energy intake. For this reason, these observations were not used in the analyses. Additionally, a preliminary sensitivity analysis and residual diagnostics  suggestThe experimental design was an incomplete block crossover design, with two of the three drink treatments given sequentially to each subject. Data were analyzed in the mixed linear models framework, using the Proc Mixed procedure in SAS (version 9.1). SubjectFor models predicting ghrelin concentration, we included 24EE, energy intake, and the interaction between 24EE and energy intake. We also considered prior day (up to 7 days) and subsequent day (up to 2 days) values for energy intake and ghrelin, and their interactions as candidate covariates. Values for up to 4 prior days for 24EE were used to predict ghrelin. For models predicting daily energy intake, we included preprandial ghrelin concentrations, 24EE, the interaction between ghrelin and 24EE, and additionally considered as candidate covariates the prior (seven days) and subsequent (two days) days for these two variables and their interactions. We explored models that included other variables and interactions, but none of those variables appeared useful. Data are presented as total intake (intake including treatment drinks) and/or selected intake (intake without treatment drinks). Values are presented as means ± SD unless indicated otherwise.Since a preliminary analysis suggested that the effect of ghrelin on energy intake was small or negligible, we conducted a power analysis to determine our ability to detect an effect of ghrelin if the effect was small. This was accomplished by Monte-Carlo simulation (creating simulated data sets based on the data we collected) and, starting with no effect of ghrelin (a true coefficient of zero for ghrelin in a regression context), determining how large the true coefficient needed to be to obtain significance for most of the simulations, at powers of 80%, 90%, and 95%, with 1000 simulations for each coefficient value. These results are most easily interpreted as how large a partial correlation between ghrelin and energy intake  would be necessary for us to detect it. We conducted this analysis for both total energy intake and breakfast energy intake .MK was responsible for statistical analysis and interpretation. DR was responsible for supervising the food intake portion of the study. WR conceived the study, and supervised the data collection and analysis. DP was responsible for ghrelin analysis, data collection, statistical analysis and manuscript preparation. All authors read and approved the final manuscript.


The information theoretic concept of mutual information provides a general framework to evaluate dependencies between variables. In the context of the clustering of genes with similar patterns of expression it has been suggested as a general quantity of similarity to extend commonly used linear measures. Since mutual information is defined in terms of discrete variables, its application to continuous data requires the use of binning procedures, which can lead to significant numerical errors for datasets of small or moderate size.In this work, we propose a method for the numerical estimation of mutual information from continuous data. We investigate the characteristic properties arising from the application of our algorithm and show that our approach outperforms commonly used algorithms: The significance, as a measure of the power of distinction from random correlation, is significantly increased. This concept is subsequently illustrated on two large-scale gene expression datasets and the results are compared to those obtained using other similarity measures.kloska@scienion.de upon request.A C++ source code of our algorithm is available for non-commercial use from The utilisation of mutual information as similarity measure enables the detection of non-linear correlations in gene expression datasets. Frequently applied linear correlation measures, which are often used on an ad-hoc basis without further justification, are thereby extended. The evaluation of the complex regulatory networks underlying molecular processes poses a major challenge to current research. With modern experimental methods in the field of gene expression, it is possible to monitor mRNA abundance for whole genomes ,2. To elTherefore, information theoretic concepts, such as mutual information, are being used to extend more conventional methods in various contexts ranging from expression -8 and DNIn this work, we discuss mutual information as a measure of similarity between variables. In the first section, we give a short introduction into the basic concepts including a brief description of the commonly used approaches for numerical estimation from continuous data. In the following section, we then present an algorithm for estimating mutual information from finite data.The properties arising from this approach are compared to previously existing algorithms. In subsequent sections, we then apply our concept to large-scale cDNA abundance datasets and determine if these datasets can be sufficiently described using linear measurements or if a significant amount of non-linear correlations are missed.A, with a finite set of M possible states {a1, a2, ... , }, the Shannon entropy H(A) is defined as [Mutual information represents a general information theoretic approach to determine the statistical dependence between variables. The concept was initially developed for discrete data. For a system, fined as p(ia) denotes the probability of the state ia. The Shannon entropy is a measure for how evenly the states of A are distributed. The entropy of system A becomes zero if the outcome of a measurement of A is completely determined to be ja, thus if p(ja) = 1 and p(ia) = 0 for all i ≠ j, whereas the entropy becomes maximal if all probabilities are equal. The joint entropy H of two systems A and B is defined analogouslywhere This leads to the relationH ≤ H(A) + H(B)     (3)A and B. Mutual information MI can be defined as [which fulfils equality only in the case of statistical independence of fined as MI = H(A) + H(B) - H ≥ 0     (4)A and B are statistically independent and increases the less statistically independent A and B are.It is zero if If mutual information is indeed to be used for the analysis of gene-expression data, the continuous experimental data need to be partitioned into discrete intervals, or bins. In the following section, we briefly review the established procedures; a description of how we have extended the basic approach will be provided in the subsequent section.p(ia) is straightforward. Many practical applications, however, supply continuous data for which the probability distributions are unknown and have to be estimated. In a widely used approach [M discrete intervals ia, i = 1... AM. For experimental data consisting of N measurements of a variable ux, u = 1... N, an indicator function Θi counts the number of data points within each bin. The probabilities are then estimated based on the relative frequencies of occurrenceIn the case of discrete data the estimation of the probabilities approach , the calwith are calculated analogously from a multivariate histogram. Additionally it has been suggested [ was suggested [For two variables the joint probabilities uggested  to adaptuggested -20. Entruggested  and showx) to the set of polynomial B-spline functions. Here, we do not provide the mathematical details for these functions since they have been discussed extensively in the literature [i, with weights given by the B-spline functions Bi,k. The spline order k determines the shape of the weight functions and thereby the number of bins each of the data points is assigned to. A spline order k = 1 corresponds to the simple binning, as described in the previous section: Each data point is assigned to exactly one bin  Determine Mx weighting coefficients for each xu from (b) Determine xu and determine p(ai) for each bin ai from(c) Sum over all H(x) according to Eq. (1)(d) Determine entropy x and y2. Calculation of joint entropy of two variables x and y, independently(a) Apply steps 1 (a) and (b) to both variables p for all Mx × My bins according to(b) Calculate joint probabilities H according to Eq. (2)(c) Calculate the joint entropy MI according to Eq. (4)3. Calculate the mutual information x = 0.0,0.2,0.4,0.6,0.8,1.0 and y = 0.8,1.0,0.6,0.4,0.0,0.2 for M = 3 bins, spline order k = 2, and the logarithm to basis two.We show the estimation with the standard binning and our approach ex-emplarily on two artificial variables p(a1) = p(a2) = p(a3) = , analogously for p(ib) due to the symmetry of data H(x) = H(y) =  = log2 3 ≈ 1.58. For the calculation of the joint probability, three of the nine two dimensional bins contain two values each p = p = p =  resulting in H = log2 3 and MI = log2 3.For both variables, each of the three histogram bins contains two values  is determined to Bi,k(2x) according to Eq. (9) (rule 1(a)). For each value xu three weighting coefficients are determined (rule 1(c)) and probabilities are calculated (rule 1(d))  = H(y) = Iog2(10) - 0.61og2(3) - 0.41og2(4) ≈ 1.57. Both, H(A) and H(B), are slightly smaller than the entropies calculated from the simple binning. The joint probabilities are p = p = 0, p = p = p = p = 0.56/6, p = p = 1.24/6, p = 1.28/6 (rule 2 (b)) resulting in H = 2.7 and MI = 0.45.The modified indicator function In the next sections, we discuss some of the properties arising from the utilisation of B-spline functions for the estimation of mutual information and compare our approach to other commonly used estimators. We support this discussion using examples for which the underlying distributions and thereby the true mutual information is known.N data points. For the simple binning approach, the mean observed mutual information can be calculated explicitly as the deviation from the true mutual informationIt has been discussed elsewhere -28,20 thN, with the slope depending on the number of bins M in accordance with Eq. (12). Figure k = 3, compared to the estimation with the simple binning (k = 1). Mutual information calculated from KDE does not show a linear behaviour but rather an asymptotic one with a linear tail for large datasets. The values are slightly increased compared to the ones from the B-spline approach. The entropy estimator  gives values comparable to the ones obtained from the B-spline approach.As can be seen for an example of artificially generated equidistributed random numbers Figure , left, mk = 1) scales with 1/N for statistically independent events [k = 3), this scaling still holds, but the average values are decreased significantly. For the KDE approach, an asymptotic run above the values from the B-spline approach is observed, again with linear tail for large datasets.  shows a linear scaling slightly below the simple binning.More importantly, a similar result also holds for the standard deviation of mutual information. As shown in Figure t events ,29. For MIdata, the average value obtained from surrogate data <MI >, and its standard deviation σsurr, the significance S can be formulated asThe interpretation of any results obtained from the application of mutual information to experimental data is based on testing to see if the calculated results are consistent with a previously chosen null hypothesis. By following the intuitive approach that the null hypothesis assumes the statistical independence of variables, mutual information is tested against a surrogate dataset, which is consistent with this null hypothesis. As discussed previously in more detail , one wayS the null hypothesis can be rejected to a certain level α depending on the underlying distribution. With increasing significance the probability of false positive associations drops.For each M = 6 bins and different spline orders k = 1... 5  to k = 2 with an increase by roughly two-fold. Using more sophisticated functions (k ≥ 3) does not further improve the significance. Similar findings have been reported in the context of kernel density estimators [k > 1 leading to smaller standard deviations σsurr.From 300 shuffled realisations of this dataset, the mean and maximum mutual information are shown with the standard deviation as error-bars. For all spline orders the null hypothesis can be rejected, in accordance with the dataset shown in Figure timators . The majk = 1 and k = 3  times higher than the ones from the B-spline approach. Strategies to simplify the integration step were proposed [ produces intermediate significances between the ones from the binning and the B-spline approach for higher bin numbers. For low bin numbers, the significances are relatively poor.Again, the significance is calculated Figure  and compproposed  but haveWe now turn to the analysis of experimentally measured gene expression data. As shown previously, the application of mutual information to large-scale expression data reveals biologically-relevant clusters of genes ,30. In tR [Among the most frequently used measures of similarity for clustering co-expressed genes are the Euclidean distance and the Pearson correlation coefficient R . If corrR X and Y we plot the tuple , R). In order to address significance, we additionally calculate all tuples from shuffled data.is expected to be fulfilled. Therefore, we calculated both, the mutual information and the Pearson correlation, for two large-scale gene expression datasets Figure . For eacS. cerevisiae for up to E1 = 300 experiments [G1 = 5345 genes. Analysis on this dataset using mutual information has been done before [The first dataset contains cDNA measurements for eriments . To avoie before ,32 on raE2 = 102 experiments on G2 = 22608 genes derived from 20 different human tissues [The second dataset contains cDNA measurements for  tissues . In contFor both datasets, tuples calculated from shuffled data Figure , and 8D After a brief introduction into the information theoretic concept of mutual information, we proposed a method for its estimation from continuous data. Within our approach, we extend the bins of the classical algorithm to polynomial B-spline functions: Data points are no longer assigned to exactly one bin but to several bins simultaneously, with weights given by the B-spline functions. By definition, the weighting coefficients for each data point automatically sum up to unity. Though our algorithm is reminiscent of kernel density estimators , it keep, we provided a systematic comparison between all these algorithms for artificially generated datasets, drawn from a known distribution. We found that mutual information, as well as its standard deviation, scales linearly with the inverse size of a dataset for the standard binning method, for the B-spline approach, and for . For the KDE approach we find an asymptotic behaviour with a linear tail for large datasets. Moreover, the discrimination of correlations from the hypothesis of statistical independence is significantly improved by extending the standard binning method to B-spline functions, as shown by a two-fold increase of the significance. Compared to KDE, the B-spline functions produce similar significances. However, due to the numerical expenses of the KDE, an application of this algorithm is limited to datasets of mod-erate size. The application of  leads to significances in-between the standard binning and the B-spline approach for reasonable bin numbers. Linear correlation measures are among the most applied measures of similarity in the literature. Often, they are used on an ad-hoc basis and it is unclear whether a considerable number of non-linear correlations are missed. Here, we asked the question whether previous analyses, based on linear correlations, sufficiently described the correlations within gene expression datasets or whether mutual information detects additional correlations that are not detected by linear measures, such as the Pearson correlation. For data that is well described by the Pearson correlation, we can give the relation of the Pearson correlation to the mutual information explicitly [To show that our approach improves the simple binning method and to compare it to KDE and the recently reported estimator plicitly . Both meplicitly ,33. We aOur findings show that the first dataset is fairly well described by the given relation of the Pearson correlation to the mutual information. No data points with high mutual information and low Pearson correlation are detected. Comparisons of genes containing outlying values, however, result in deviations with low mutual information and high Pearson correlation. From this, it follows that previous analyses on this dataset, based on Pearson correlation, did not miss any non-linear correlations. This presents an important finding since it is by all means supposable that the regulations inherent in the genetic network under consideration might show more complex behaviour than the observed linear ones. Even for one of the largest expression datasets at hand, insufficient data might complicate the detection of such complex patterns of regulation. Alternatively, the biological mechanisms which underlay the regulatory networks might not lead to non-linear correlations. It also has to be considered that the experimental methods applied for the generation of this dataset may make non-linear correlations difficult to detect. The second dataset, in contrast, reveals highly significant tuples with high mutual information and low Pearson correlation. Detailed gene-gene plots of such tuples show that the expression values of the contributing genes fall into groups of experimental conditions. Without attempting to draw conclusions about the biological context of such clusters here, they might reflect interesting situations worth to be analysed in detail.Most of the manuscript text was written by CD and edited by all authors. CD carried out the calculations and produced the figures. RS strongly contributed to the theoretical background of entropy and mutual information.The implementation of the C++ program was carried out by SK. JS and SK supervised this work. All authors read and approved the final manuscript.


Wasting and micronutrient malnutrition have not been well characterized in adults with pulmonary tuberculosis. We hypothesized that micronutrient malnutrition is associated with wasting and higher plasma human immunodeficiency virus (HIV) load in adults with pulmonary tuberculosis.In a cross-sectional study involving 579 HIV-positive and 222 HIV-negative adults with pulmonary tuberculosis in Zomba, Malawi, anthropometry, plasma HIV load and plasma micronutrient concentrations  were measured. The risk of micronutrient deficiencies was examined at different severity levels of wasting.Body mass index (BMI), plasma retinol, carotenoid and selenium concentrations significantly decreased by increasing tertile of plasma HIV load. There were no significant differences in plasma micronutrient concentrations between HIV-negative individuals and HIV-positive individuals who were in the lowest tertile of plasma HIV load. Plasma vitamin A concentrations <0.70 μmol/L occurred in 61%, and zinc and selenium deficiency occurred in 85% and 87% respectively. Wasting, defined as BMI<18.5 was present in 59% of study participants and was independently associated with a higher risk of low carotenoids, and vitamin A and selenium deficiency. Severe wasting, defined as BMI<16.0 showed the strongest associations with deficiencies in vitamin A, selenium and plasma carotenoids.These data demonstrate that wasting and higher HIV load in pulmonary tuberculosis are associated with micronutrient malnutrition. Mycobacterium tuberculosis, and the majority live in less developed countries where human immunodeficiency virus (HIV) infection is spreading rapidly. The World Health Organization (WHO) estimates that the number of new cases of tuberculosis and the proportion with coexisting HIV infection will continue to increase [6 [Approximately one-third of the world's population is infected with increase . Immunosincrease -5. Microincrease -17 and iincrease . Severalincrease -12, thiaincrease , vitaminrease [6 , folate rease [6 , vitaminrease [6 , and zinrease [6 . Poor serease [6 , but selThe relationship between severity of HIV disease and micronutrient malnutrition needs further elucidation. Such information would help identify subgroups that might benefit the most from nutritional interventions. Plasma HIV load was used as an indicator of severity of HIV disease, as HIV load tends to be higher in more active HIV disease. We hypothesized that wasting in pulmonary tuberculosis is associated with micronutrient malnutrition and that HIV-infected adults with pulmonary tuberculosis who have more active HIV disease, as reflected by higher HIV load, also have more severe micronutrient malnutrition. To test these hypotheses, we conducted a cross-sectional study to examine the relationship between wasting and micronutrient malnutrition in HIV-positive and HIV-negative adults with pulmonary tuberculosis in Zomba, Malawi.+ lymphocyte counts were not conducted due to limited resources. None of the participants were taking antiretroviral treatment.The study population consisted of adults who presented with new sputum-positive pulmonary tuberculosis in Zomba Central Hospital between July 1999 and April 2003. Subjects were offered HIV testing and were screened for HIV antibodies after signing a written informed consent form. All subjects were given appropriate pre- and post-test HIV counseling. Subjects commenced treatment after enrollment and received standard short course chemotherapy for tuberculosis as per guidelines of the Malawi National Tuberculosis Program . Adults The protocol was approved by the institutional review boards at the Johns Hopkins School of Medicine  and the College of Medicine, University of Malawi , with final approval by the Office for Protection from Research Risk of the National Institutes of Health.2.Body weight was determined to the nearest 0.1 kg using an adult balance , and standing height was determined to the nearest cm. Body mass index (BMI) was calculated as body weight/heightA venous blood sample was collected by venipuncture . Blood samples were shielded from bright light and immediately aliquoted and stored in cryotubes at -70°C. α-carotene, β-carotene, β-cryptoxanthin, lycopene, lutein, zeaxanthin, retinol, and α-tocopherol concentrations were measured in 100 uL of plasma by high performance liquid chromatography using a modified method from the Nutrition Laboratory, Inorganic Toxicology and Nutrition Branch Division of Laboratory Sciences, National Center of Environmental Health, Centers of Disease Control and Prevention  . Total pt-tests and nonparametric Mann-Whitney U-tests. Univariate analysis of variance was used to test for linear trends across categories of plasma HIV load and BMI.Data and statistical analysis were conducted using SAS 8.01  and SPSS 9.0 . Comparisons between groups were made using HIV load was categorized into tertiles. HIV negative subjects were assigned a fourth category of HIV load (category 0) when groups were merged for analysis. Nutritional status was assessed in adults with pulmonary tuberculosis with and without HIV co-infection. Subjects were separated into groups according to their extent of wasting. Mild wasting was defined as BMI 17.0–18.49, moderate wasting as BMI 16.0–16.99, and severe wasting as BMI <16.0, conform the WHO strata for BMI grading of severity of malnutrition .Plasma retinol <0.70 μmol/L was considered consistent with vitamin A deficiency . VitaminP < 0.01 was used in this study.To examine the risk of micronutrient deficiencies at different severity level of wasting, logistic regression models were fitted with retinol <0.70, α-tocopherol <11.6, zinc <11.5, selenium <0.89, and the lowest quartile of total carotenoids as the outcome variable. Multivariate logistic regression models were conducted to adjust for sex, age and HIV load. A significance level of The study population consisted of 579 HIV-positive and 222 HIV-negative adults with sputum-positive pulmonary tuberculosis. Among the total study population, 66% (232/352) of male and 77% (347/449) of female participants were HIV-positive. The mean age among all subjects was 33 years (range 18–59 years). The majority of subjects were wasted, as 59% of subjects had a BMI <18.5, 32% of subjects had a BMI <17.0, and 17% of all subjects were severely wasted as defined by BMI<16.0. Plasma retinol concentrations <0.70 μmol/L occurred in 61% of all subjects. Vitamin E, zinc, and selenium deficiency occurred in 13%, 85% and 87% respectively. Table Table Figures P = 0.004). Total carotenoid concentrations significantly decreased with the increase of plasma HIV load among non-wasted, mildly wasted, moderately wasted and severely wasted adults . Selenium concentrations decreased significantly with the increase of plasma HIV load among non-wasted and severely wasted adults with pulmonary tuberculosis . Among the HIV negative adults and those in the 1st and 2nd tertile of HIV load, plasma retinol, total carotenoids and selenium concentrations significantly decrease with the increasing severity of wasting. Among those in the 3rd tertile of HIV load, only plasma retinol concentrations significantly decreased with the increasing severity of wasting. This trend did not reach significance for plasma carotenoid and selenium concentrations.Plasma retinol concentrations significantly decreased with the increase of plasma HIV load among non-wasted adults with pulmonary tuberculosis  declare that they have no competing interests.Overall guidance and initial study design was provided by RS. MvL has been in charge of the collection and analysis of data and writing of the manuscript. Provision of advice was given by AH, JK, EZ, TC and TT. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:


Plant genomes contain various kinds of repetitive sequences such as transposable elements, microsatellites, tandem repeats and virus-like sequences. Most of them, with the exception of virus-like sequences, do not allow us to trace their origins nor to follow the process of their integration into the host genome. Recent discoveries of virus-like sequences in plant genomes led us to set the objective of elucidating the origin of the repetitive sequences. Endogenous rice tungro bacilliform virus (RTBV)-like sequences (ERTBVs) have been found throughout the rice genome. Here, we reconstructed putative virus structures from RTBV-like sequences in the rice genome and characterized to understand evolutionary implication, integration manner and involvements of endogenous virus segments in the corresponding disease response.Oryza AA-genome species vary in the ERTBV copy number. The species carrying low-copy-number of ERTBV segments have been reported to be extremely susceptible to RTBV. The DNA methylation state of the ERTBV sequences was correlated with their copy number in the genome.We have collected ERTBVs from the rice genomes. They contain rearranged structures and no intact ORFs. The identified ERTBV segments were shown to be phylogenetically divided into three clusters. For each phylogenetic cluster, we were able to make a consensus alignment for a circular virus-like structure carrying two complete ORFs. Comparisons of DNA and amino acid sequences suggested the closely relationship between ERTBV and RTBV. The Oryza AA-genome species allowed us to speculate a possible role of endogenous virus segments against its related disease.These ERTBV segments are unlikely to have functional potential as a virus. However, these sequences facilitate to establish putative virus that provided information underlying virus integration and evolutionary relationship with existing virus. Comparison of ERTBV among the  AL1 gene, have been found in the genomes of tobacco and its related species [The virus-like sequences that have been found in plant genomes are divided into two groups of plant viruses, single-stranded DNA geminivirus and double-stranded DNA pararetroviruses. The geminivirus segments, including the viral replication origin and the adjacent  species . Pararet species ,4 banana species -7 and to species -10. Comp species . The end species , tobacco species  and petu species , but theOryza glaberrima and O. barthii, showed much severer systemic necrosis compared to the other rice species present in South and Southeast Asia after inoculation of both RTBV and RTSV.In the rice genome, pararetrovirus-like sequences that are similar to rice tungro bacilliform virus (RTBV) have also been found -14. In SOryza species. Asian species have more ERTBV segments than the species originated from the other regions where RTBV is not distributed. The results obtained advance our understanding of the manner of integration of authentic pararetrovirus into the host genome, evolutionary implication of the integrated virus and possible involvement of endogenous virus segments in the corresponding disease resistance.Here, we have characterized RTBV-like sequences in the Japonica (cv. Nipponbare) genome. These sequences, denoted endogenous RTBV-like sequences (ERTBVs), were highly rearranged and dispersed throughout the rice genome. Sequences of the putative viruses for ERTBV were reconstructed from the dispersed segment in the genome. Copy numbers of ERTBV segments are shown to vary among AA-genome EcoRI-digested fragments from both Japonica and Indica genomic DNAs [Previously, we found a repetitive sequence near the rice waxy gene, which is partially homologous to the rice tungro bacilliform virus (RTBV) genome ,17. A prmic DNAs . In the We sorted out the 29 ERTBV segments using the sequences homologous to the RT gene, which is encoded in the end of ORF 3. Similarity analyses based on the RT gene grouped them into three clusters Figure , suggestThe longest ORF in RTBV, ORF 3 encodes the movement protein (MP), coat protein (CP), asparatic protease (PR), RT and RNase H (RH) in the single polycistronic mRNA . ORF 3 mTo evaluate the genetic relationship of ERTBVs and RTBV, RT amino acid sequences from the viruses belonging to Caulimoviridae, were compared. Using the PHYLIP package program , the phyOryza AA-genome species  using the RTBV-like sequence near the es Table . Indica es Table , indicatExcept for the sequences examined here, we could not find any other RTBV-like sequences in database searches using RTBV nor each of ERTBVs sequences as queries. Therefore, a fourth cluster of the RTBV-like sequences is unlikely to be present. The integration of ERTBV or its derived segments after the differentiation of Japonica and Indica cultivars thus seems to have not occurred.Oryza species contain ERTBV in their genomes, Southern blotting experiments were performed with 14 lines of the Oryza AA-genome species. A PCR fragment of the 7.4-kb ERTBV sequence from chromosome 10 was prepared as a probe, and Figure O. sativa and O. rufipogon (lanes 1–8) has about 50 bands, some of which showed common sizes among the lines. An Australian rice  and two O. longistaminata accessions showed middle-copy numbers (approximately 10–20) of the ERTBV segments. In O. glaberrima, O. barthii and O. glumaepatula accessions, only a few bands hybridized to ERTBV; this may also be caused by divergence of the ERTBV sequence and not only due to lower copy numbers of the sequence. Especially in These three species derived from Africa and Latin America were genetically distinct from Asian rice species, but another African rice species,O. longistaminata, was considered to have undergone introgression with Asian species [To test whether other  species . Africa Oryza species with various copy numbers. The methylation states were analyzed by Southern blotting method using a methylation-sensitive enzyme, HpaII, and a partially methylation-sensitive isoschizomer, MspI. The 7.4-kb probe was employed to examine the methylation states of the ERTBV in the Oryza AA-genome species. The blotting patterns of O. sativa (Shimokita), O. rufipogon (W1954), O. longistaminata (W1034), O. meridionalis (W1625), O. barthii (W1592), O. glumaepatula (W1185) showed differences of methylation state depending on the ERTBV copy number -like sequences [Brassica genome show strong affinity for S/MAR-like sequences [Our data suggest that the putative virus for ERTBVs has been integrated into the  tobacco  and hepa tobacco  were als [et al.  proposed [et al. . This faequences ,25. SINEequences . SimilarOryza AA-genome species before differentiation of Japonica and Indica. Considering this evolutionary period, two possible relationships of RTBV and ERTBV might be predicted: one is that the virus that was to become ERTBV was a direct progenitor of RTBV, and the other is that both were differentially branched from a common ancestral virus. Because the evolution of virus genomes is generally much faster than that of plant genomes, we cannot compare them equivalently. In fact, rapid evolution of RTBV was inferred from high level of genetic diversity of RTBV field isolates [Although the genomes of ERTBV and RTBV are structurally similar, particularly in the longest ORF, some regions are markedly different. Phylogenetically, both are apparently closely related viruses. One important question is whether or not ERTBV is a cognate of RTBV. ERTBVs existed in the isolates . Even inisolates . Particuisolates . If ERTBisolates  and is tIn the other case, in which RTBV and the virus for ERTBV are phylogenetically located on different branches, the progeny of the virus corresponding to ERTBV might have vanished or may be hidden in a small population. Even though RTBV is not in the case a direct progeny of the virus corresponding to ERTBV, their structural similarity and parallel distribution lead us to consider their common ancestral origin.O. glaberrima and O. barthii showed severe systemic necrosis within 4 weeks after infection of RTBV and rice tungro spherical virus (RTSV). O. longistaminata, which possesses ERTBV fragments in its genome, showed disease symptoms like those of the Asian species, and some of the accessions were utilized for obtaining the resistance gene [et al. [Plant virus-like sequences have been found in several plant genomes such as banana, tobacco and petunia ,31. So fnce gene . Interes [et al.  proposedOryza speciation. ERTBV integrations likely occurred when the virus was in the replication process, and were preferentially targeted to AT-repeat sequences. The closely relationship between ERTBV and RTBV were proven by comparisons of the DNA and amino acid sequences. The Oryza AA-genome species originated from RTBV-distributed regions appeared to contain higher copy numbers of ERTBV segments. The methylation state of the ERTBV sequences was correlated with their copy number in the genome. The results obtained allowed us to speculate a possible relationship between RTBV disease resistance and the copy number and/or DNA methylation of ERTBV in the Oryza AA-genome species.The rice genome contains more than 30 of RTBV-like sequence (ERTBVs) which were unlikely to have functional potential as a virus, while we were able to assemble putative virus forms from these sequences. The phylogenetic analysis showed that at least three times integrations of authentic ERTBVs occurred during Sequences of ERTBV in Japonica (cv. Nipponbare) were mined with rice blast search queries  against waxy locus was used as probe [We attempted to search for ERTBV sequences in the Indica (cv. 93–11) genome database, however, the homologous sequences found through the search had insufficient length for designation as ERTBV segments. Indica ERTBV sequences were isolated from the EMBL 3 genomic library constructed with IR36 strain (FL1041j), which was purchased from Clontech . For screening, a 3.5-kb ERTBV fragment about 50-kb upstream of the as probe . Three cThe nucleotide sequences and amino acids sequences inferred from the reverse transcriptase (RT) gene were aligned using the CLUSTAL W program  from theOryza AA-genome species including 14 strains , and hybridized using the Alk Phos Direct Southern hybridization kit (Amersham Life Science). For the probe, nearly the full length of the 7.4-kb ERTBV fragment  was amplified by PCR with the primer combination of 5'GAACTACAACTAGATATGAACGGGGATA3'+5'CACAACTATTCTTAGTGCTGAATTCACTT3'. The membranes were washed twice under the standard Alk Phos conditions with 0.5 M NaCl at 42°C for 20 minutes. To test the methylation state of ERTBV, Southern blotting analysis was carried out using the C-methylation-sensitive enzyme HpaII and the partially sensitive enzyme MspI (isoschizomer of HpaII). The same probe mentioned above was prepared using a PCR-based labeling system with the PCR DIG labeling mix (Roche). To verify that complete digestion was achieved by the enzymes, a chloroplast DNA fragment, the 5.2-kb Sma-8 fragment of buckwheat [Total genomic DNA was isolated from 7 ns Table . EcoRI-duckwheat , was useThe sequences containing Indica ERTBV have been deposited in DDBJ: accession nos. AB124591, AB124592 and AB124593. The reconstructed sequences for the authentic ERTBV viruses, ERTBV-A, -B and -C were deposited in DDBJ: accession nos. BR000029, BR000030 and BR000031, respectively.MoKu and MaKa carried out the molecular genetic studies and participated in the sequence alignments. HN participated in the sequence alignment and performed the phylogenetic analysis. IU participated in the design of the study. YK conceived of the study and drafted the manuscript. YK and YS participated in its design and coordination. All authors read and approved the final manuscript.


Microvascular alterations contribute to the development of stroke and vascular dementia. The goal of this study was to evaluate age and hypertension related changes of the basal lamina in cerebral microvessels of individuals, who died from non-cerebral causes.We examined 27 human brains: 11 young and 16 old patients. Old patients were divided into two subgroups, those with hypertension (n = 8) and those without hypertension (n = 8). Basal lamina changes of the cerebral microvessels were determined in the putamen using antibodies against collagen type IV and by quantitative analysis of vessel number, total stained area of collagen, thickness of the vessel wall and lumen, and relative staining intensity using immunofluorescence. The total number of collagen positive vessels per microscopic field was reduced in old compared to young subjects . The relative collagen content per vessel  and the relative collagen intensity  shown by immunofluorescence were higher in the older compared to the younger patients with a consecutive reduction of the lumen / wall ratio . No differences were observed for these parameters between old hypertensive and non-hypertensive patients.The present data show age-related changes of the cerebral microvessels in sections of human putamen for the first time. Due to the accumulation of collagen, microvessels thicken and show a reduction in their lumen. Besides this, the number of vessels decreases. These findings might represent a precondition for the development of vascular cognitive impairment. However, hypertension was not proven to modulate these changes. Aging is associated with a deterioration of cognitive function including a decrease in the ability to process and store new information . ProcessAlthough changes of cerebral vessels with aging and hypertension have been reported, there are presently no consistent data on the microvascular basal lamina. The aim of our study was an evaluation of age-related changes on cerebral microvessels and on the vascular extracellular matrix in humans and the possible impact of chronic hypertension using several different immunohistochemical methods for the detection of collagen type IV.Old patients (OP) showed 12.0 +/- 0.6 vessels per microscopic field, and young patients (YP) exhibited 15.1 ± 1.2 vessels per microscopic field (p = 0.02). The total area of collagen type IV positive vessels did not differ between OP and YP . Therefore the calculated content of collagen type IV per vessel was higher in OP than in YP . No differences were observed between old non hypertensive patients (ONHP) and old hypertensive patients (OHP) .Microvessel wall thickness, inner lumen and ratio lumen/wall thickness were statistically different in OP vs. YP. The vessel wall was thicker in old than in young patients . In addition the vessel lumen was reduced in the old group compared to the young group . Again, the intensity in non-hypertensive old persons showed no difference compared to hypertensive old patients , and old patients . The old patients were divided into two subgroups, those without a history of hypertension , and those with a history of hypertension . There were no significant differences in age between OHP and ONHP and in sex between YP and OP, as well as between ONHP and OHP  in 0.01 N HCl for one hour. Collagen IV-positive vessels were stained with a monoclonal mouse anti-collagen-IV antibody . Each section was incubated with 150 μl of the primary antibody solution (at a concentration of 1:200) for two hours at 37°C followed by incubation with biotinylated secondary antibody against mouse IgG for 30 minutes at 37°C (Vector Laboratories). Vectastain ABC reagent was added for 30 minutes at 37°C. Chromogen (AEC Kit Biomeda Corp.) was used to develop the peroxidase signal. Negative and positive controls were routinely performed in each staining experiment. The same procedure was used for immunofluorescence staining. Instead of using the Vectastain ABC kit containing avidin, avidin marked FITC  was added for 30 minutes at a dilution of 1:100.The number of peroxidase stained vessels was determined with the aid of a computerized video imaging system at a magnification of ×100 . Only vessels smaller than 30 μm were included. Total area of collagen IV positive vessels in the sections was analyzed using the same system. Results are presented in arbitrary units. To obtain the relative amount of collagen type IV per vessel the area of collagen type IV was divided by the number of stained vessels per microscopic field ([collagen type IV/microscopic field]/[vessels/microscopic field]) The size of the observed microscopic field was 150 × 200 μm.To estimate microvessel hypertrophy, the ratio between the diameter of vessel lumen and vessel wall, respectively, was calculated semiquantitatively with the help of a second computerized video imaging system . Twenty entire cross-sectional microvessels from the putamen were randomly selected at a magnification of ×400. To calculate the wall to lumen ratio, average distances of vessel wall and vessel lumen were selected.z plane (10 scans per 1 μm), and a summed image was calculated. Also, a summed image was obtained from the background area to normalize the local intensity to the background. The normalized intensity is expressed as mean ± SEM for each microvessel using a scale from 0 to 255 arbitrary units (U). The technique was adopted from Hamann et al [Fluorescence intensity measurements of microvessel-associated FITC anti-mouse IgG against the anti-collagen antibody were performed with confocal laser scanning microscopy . All measurements were performed with the same pinhole size, brightness and contrast, zoom, and laser time. Each vessel was scanned in the nn et al . Twenty Data are presented as mean +/- standard error of mean. Statistical evaluations were performed using t-test.OU carried out the immunohistochemical experiments, ML performed the statistical analysis and drafted the manuscript. JH participated in the design of the study and collected the brain specimens. AD participated in the study design. GFH supervised the thesis, and participated in its design and coordination.