Other 3-nitrobenzoic acid derivatives showed a peculiar CA inhibition profile with a notable potency towards hCA IX.Selective inhibition of tumor-associated carbonic anhydrase  isoforms IX and XII is a crucial prerequisite to develop successful anticancer therapeutics. Herein, we confirmed the efficacy of the 3-nitrobenzoic acid substructure in the design of potent and selective carboxylic acid derivatives as CAs inhibitors. Compound  To date, 16 isoforms have been characterized in humans and are involved in many physiological and physio-pathological processes, like the regulation of pH and acid–base homeostasis, CO2 elimination, bone resorption, electrolyte secretion, gluconeogenesis, ureagenesis, and tumorigenesis [Carbonic Anhydrases  are ubiquitous Zn-dependent metalloenzymes that catalyze the interconversion of carbon dioxide  used in the clinical practice or in clinical development for the treatment of glaucoma , epilepsin silico target fishing approach to recycle a false-positive hit Ki values were significantly higher with respect to the reference compounds AAZ and GV2-20. The drop of inhibitory potency compared to GV2-20 is particularly marked against hCA II, where the most potent inhibitor of the series, namely compound 1, has a Ki around 54-fold higher than that of AAZ and around 10-fold higher than that of GV2-20. Sub-micromolar inhibition of hCA I was observed only for compounds 1 and 8, bearing respectively a 1,3,4-oxadiazole and a 1,3,4-thiadiazole moiety in the tail, while 1 was the only hCA I inhibitor with sub-micromolar Ki. Modifications to the heterocycle such as the tetrazole in 17 proved ineffective for hCA II inhibition. Other tested compounds showed Ki in the micromolar range, or proved inactive at all against these cytosolic hCA isoforms.All tested compound are poorly active against cytosolic hCA I and hCA II, as (ii)Similar to the reference inhibitor GV2-20 but different from AAZ, tested compounds proved generally inactive also against hCA IV. (iii)19, in which the carboxylic acid is replaced by the nitrile function thus opening new venues for the design of hCA VA-specific inhibitors. Other compounds with a low Ki against this isoform are 10, 12, 13, 15, and 16 that also show a certain degree of specificity for hCA VA particularly compared to cytosolic hCA I and hCA II, hCA IV, and hCA VII. Compared to GV2-20, most of tested compounds showed a significant improvement of hCA VA inhibition. Notably, the most potent hCA VA inhibitor of the series was (iv)1 was the most potent inhibitor of hCA VII, even though the chemically-related compound 8 proved inactive. Compared to GV2-20, compound 1 showed around 37-fold drop of inhibitory potency against hCA VII. Moreover, 5, 13, and 19 that are structurally-related to GV2-20 proved inactive. This suggests that subtle modifications to GV2-20 scaffold have a dramatic impact on hCA VII inhibition, and that the 3,5-dinitrobenzoic acid is required to target this hCA isoform. Most of tested compound were poorly active against hCA VII that, in contrast, was efficiently inhibited by GV2-20 and AAZ. As already observed for hCA I and hCA II, compound (v)6–8, 10, 14–17, whereas none of them was able to overtake AAZ or GV2-20 against hCA XII. The most potent inhibitor of hCA IX was 10, which also inhibited hCA XII in the low nanomolar range. Unique among other is the behavior of compounds 6 and 7, which are low nanomolar inhibitors of CA IX with potency comparable to AAZ, although they are not active against hCA XII. Of note, 7 is the lowest MW compound of the test-set, thus becoming the selective hCA IX inhibitor endowed with the highest ligand efficiency identified in this work [1 showed the weakest specificity for tumor-associated hCAs, whereas 10 emerged as the most selective one. Tumor-associated isoforms hCA IX and hCA XII were potently inhibited by most of tested compounds. It is worth noting that some compounds showed stronger or at least similar inhibition of hCA IX compared to the reference compounds AAZ and GV2-20, see for example his work . Therefo(vi)18 that is deprived of the nitro group in the head portion is poorly active in most hCA isoforms, with the only exception of hCA IX, for which it showed a Ki of 72.8 nM. Based on the comparison with the hCAs inhibition profile of GV2-20, we may speculate that at least one nitro group in the head portion is essential for the efficient inhibition of most hCAs. Compound 19 that bears a nitrile instead of carboxylic acid group, shares a similar hCAs inhibition profile as 18 with the exception of a stronger inhibition of hCA VA (Ki = 92.5 nM).The GV2-20 derivative Further analysis of 1 as an interesting pan-hCAs inhibitor with a moderate selectivity for tumor-associated hCA IX and hCA XII with respect to cytosolic hCA II and hCA VA, while 10 proved to be an attractive low MW anticancer lead candidate endowed with potent and selective inhibition of hCA IX and hCA XII. The highest efficiency ligand 7 emerged as selective inhibitor of hCA IX, particularly with respect to hCA XII, which paves the way to the design of tool compounds for isoform-specific modulation of hCA IX. The 3-nitrobenzoic acid substructure was confirmed as privileged head for the design of potent and selective carboxylic acid CAIs. In summary, SAR analysis highlighted compound 1, 7, 10, 15, and 19 were docked against the crystallographic structure of hCA IX, while 1, 7, and 10 were docked towards hCA XII. Indeed, 1 was an interesting pan-CAs inhibitor with a moderate specificity for hCA IX and hCA XII; 7 was the lowest MW compound showing very potent inhibition of hCA IX with no effects on hCA XII; 10 was the most potent inhibitor of hCA IX showing a good inhibition of hCA XII and the highest selectivity with respect to the cytosolic and widely expressed hCA II; 15 and 19 showed a remarkable potency and selectivity against hCA IX and hCA XII, and were notably active against hCA VA for which no crystal structures are available to date. Since the binding mode of carboxylic acid CAIs is highly variable , and ourFor the sake of clarity, in our study we addressed three different binding modes hypothesis: (i) direct binding of the carboxylic acid group to the catalytic Zn(II) ion; (ii) binding of carboxylic acid group to the Zn(II)-coordinated water molecule; and (iii) binding “out of the binding site”. Results discussed thereafter refer to the most reliable binding poses that were generated according to binding mode hypothesis (ii), namely carboxylic acid CAIs that bind the catalytic Zn(II) ion through a bridging water molecule. Docking poses were then refined by energy minimization in explicit water solvent. 1, 7, 10, and 15 binds the Zn(II)-coordinated water molecule while compound 19, in which the carboxylic acid is replaced by a nitrile group, does not, which may account for lower inhibition potency of this compound against hCA IX. Besides anchoring the Zn(II)-coordinated water molecule, the carboxylic acid group of compounds 1, 7, and 15 performs additional H-bonds with the side chain of Thr-199 and Thr-200  ion by means of a bridging water molecule  ion, nor the Zn(II)-coordinated water molecule, as observed instead for other compounds  coordination system, which strongly corroborates the binding mode predicted by molecular modeling. Of particular interest is the analysis of the binding-mode of compound ompounds A,B. The ions see . Moreoveoumarins . As show7, hCA XII, and hCA IX . Molecules were subsequently filtered using the Filter program from OpenEye (version 2.5.1.4) ,57 by usCompounds selected for CAs inhibition assays were purchased from MolPort and tested as received.Molecular docking of most interesting compounds was performed with GOLD program (version 5.2.1) . The cryLigands protonation state was assigned by fixpka application from QUACPAC  . FinallyThe best docking-based binding pose of each compound was subsequently relaxed in explicit water solvent with AMBER 16 . The AMBThe electrostatic surface potential of hCA IX and hCA XII was calculated with APBS 1.4.1 with the nonlinear Poisson−Boltzmann equation using the default parameters . 7 and hCAs were generated with PyMol [Figures describing ligands binding modes and electrostatic surfaces of compound th PyMol .7 was calculated using the Schrödinger’s Maestro Suite for molecular modelling by solving the Poisson−Boltzmann equation and using the partial charges of the molecules 7 [The electrostatic surface potential for compound ecules 7 .2 hydration activity [2SO4 (for maintaining constant the ionic strength), following the initial rates of the CA-catalysed CO2 hydration reaction for a period of 10–100 s. The CO2 concentrations ranged from 1.7 to 17 mM for the determination of the kinetic parameters and inhibition constants. For each inhibitor, at least six traces of the initial 5–10% of the reaction have been used for determining the initial velocity. The uncatalysed rates were determined in the same manner and subtracted from the total observed rates. Stock solutions of inhibitor (0.1 mM) were prepared in distilled–deionized water, and dilutions up to 0.01 nM were done thereafter with the assay buffer. Inhibitor and enzyme solutions were preincubated together for 15 min at room temperature prior to assay, in order to allow for the formation of the enzyme–inhibitor complex. The inhibition constants were obtained by non-linear least-squares methods using PRISM 3 and the Cheng–Prusoff equation, as reported earlier [in-house as reported earlier [An applied photophysics stopped-flow instrument has been used for assaying the CA catalysed COactivity . Phenol  earlier ,70,71, a earlier ,73.Nowadays, the development of isoform-selective CAIs has become a major focus in the discovery of innovative therapies for a wide range of human diseases.10 emerged as the most promising inhibitor of the tumor-associated isoforms hCA IX and hCA XII with an impressive hCA II/hCA IX and hCA II/hCA XII selectivity index, while compound 7 was a low nanomolar and highly selective inhibitor of hCA IX with the highest ligand efficiency. Finally, the 3-nitrobenzoic acid was confirmed as a good scaffold for the design of CAs inhibitors. Starting from the pan-hCAs inhibitor GV2-20, in this study we performed a tail approach in order to improve potency and selectivity of inhibition among hCAs. Notably, almost all tested compounds showed potent inhibition of hCA IX and increased selectivity with respect to other isoforms such as the widely expressed cytosolic hCA I and hCA II. Of note, compound


Various scales have been used to perform a quick and first level nutritional assessment, and the MNA is one of the most used and recommended by experts in the elderly in all areas. This scale has a short form, the MNA-SF, revised and validated in 2009, which has two versions: the BMI-MNA-SF contains the first six items of the full scale including Body Mass Index while the CC-MNA-SF includes Calf Circumference instead of BMI.To evaluate the predictive ability for nutritional status of the two versions of the MNA-SF against the MNA in free-living elderly in the province of Valencia.Cross-sectional study of 660 free-living elderly in the province of Valencia selected in 12 community centres using stratified sampling by blocks. Inclusion criteria: being aged 65 or over, living at home, having functional autonomy, residing in the province of study for more than one year, regularly attending community centres and voluntarily wanting to take part.ρ = 0.78p < 0.001; CC-MNA-SF: ρ = 0.78p < 0.001). In addition we obtained a very high correlation between the two MNA-SFs (ρ = 0.96p < 0.001). We evaluated the agreement between the full MNA and the MNA-SFs classification in three nutritional categories  with Cohen’s kappa coefficients . These values indicate moderate agreement with the full MNA. There is very good agreement between the BMI-MNA-SF and CC-MNA-SF (κ = 0.88p < 0.001). In order to determine the ability of both MNA-SFs to identify subjects not requiring any nutritional intervention, we considered the dichotomised categorisation of the full MNA and the MNA-SFs as “normal nutritional status” vs. “malnutrition and risk of malnutrition” Areas under the ROC curves using MNA as the gold standard indicate moderately high prognostic accuracy . Both versions of the MNA-SF showed similar sensitivity, specificity and diagnostic effectiveness .Of the 660 subjects studied, 319 were men (48.3%) and 341 (51.7%) women with a mean age of 74.3 years (SD = 6.6). In terms of nutritional assessment, using the BMI-MNA-SF and the CC-MNA-SF we found that 26.5% and 26.2% were at risk of malnutrition and 0.9% and 1.5% were malnourished respectively. With the full MNA, 23.3% were at risk of malnutrition. Spearman’s rank correlation coefficients indicate a high association between the full MNA score and the MNA-SFs scores (BMI-MNA-SF: In its two versions the MNA-SF presents useful predictive ability against the MNA. The advantage of the CC-MNA-SF is that using it requires fewer resources and less time in primary care, although always the characteristics of the population must take into account to make the right decision based on the MNA-SF scales. In recent years we have seen unprecedented global demographic change in the ageing of the world population. This social phenomenon will continue to grow and experts forecast that by 2050 the elderly will account for over a third of the total population of some developed countries. At 34.4%, Spain will be the fourth country in the world with the highest percentage of people over 65 .The elderly are a highly vulnerable group susceptible to nutritional difficulties  which ifCurrently it is estimated that the majority of the elderly (96.2%) are free living and want to remain at their home or at a relative’s home as long as possible . The 201Given the forcast of experts this population group is set to increase, so it would be advisable and effective to set up health programmes in primary healthcare which would enable early identification of nutritional risk, thus avoiding malnutrition and its possible consequences. Only through adequate, quick and inexpensive nutritional assessment can some nutritional problems be resolved or at least be alleviated .Various scales have been used to perform a quick and first level nutritional assessment, and the Mini Nutritional Assessment (MNA) is one of the most used and recommended by experts in the elderly in all areas. One of its main advantages is the identification of malnutrition risk before clinical alterations onset and thus does not require laboratory tests. Since 1994 and after 20 years of development and use, the MNA scale has become established as one of the most commonly used tools by researchers for first-level nutritional status assessment of the elderly, irrespective of where they live. As it is easy to use, inexpensive and reliable it has become part of the clinical practice of health professionals and a valid and essential element in comprehensive geriatric assessment .This internationally validated and used scale has a short/screening version, the Mini Nutritional Assessment Short Form (MNA-SF), which was developed and validated in 2001 . With thRecently, the MNA scale has been used as the gold standard to develop other short nutritional assessment questionnaires for free-living elderly. Some authors  have comWe designed a cross-sectional descriptive study including 660 people of both sexes aged 65 or over, free-living and residing in the province of Valencia. The study was conducted between October 2008 and November 2009. We collected data in twelve community centres for the elderly across the province of Valencia, which were selected using stratified sampling by blocks to ensure the various geographical settings of the province were represented. We chose a sample size that would ensure an estimate of the proportion of risk or malnutrition with an accuracy of 5% and a power of 80% for the tests performed, considering an estimated population prevalence of 35%.Inclusion criteria for participating in the study were being aged 65 or over, living at home, having resided in the province of Valencia for over one year, regularly attending community centres and voluntarily wanting to take part in the study.National Institute of Statistics (INE) data show that in 2009, people aged 65 and over registered in the province of Valencia accounted for 15.9% of its total population. 95% of them lived in the community (in free-living) and were regular users of socio-health centres. Valencia’s health system is free and universal and the province of study has 11 health departments with 142 health centres and 281 auxiliary offices distributed throughout its territory. They draw up health plans for the elderly to improve their healthy habits and lifestyles .We used three scales, the full MNA and its two short versions (BMI-MNA-SF and CC-MNA-SF), for the nutritional assessment of the participants. We used the full MNA as the gold standard for nutritional assessment.The full MNA scale has 18 items, provides a maximum possible overall assessment of 30 points and its nutritional categories are normal nutritional status (24–30 points), risk of malnutrition (17–23.5 points) and malnutrition (<17 points) . The MNAThe data to complete the test was obtained from personal interviews conducted in the community centres by nurses who had been trained beforehand for that purpose. The instructions in the guide to completing the MNA test  were folCC and MAC were measured with a SECA 201 tape measure (range 0–205 cm and 1 mm accuracy). MAC was measured in the non-dominant arm at the midpoint between the acromion and the olecranon with the arm hanging down and parallel to the body. CC was measured in the calf area where the circumference is greatest with the person sitting down, resting their foot on the floor and their knee bent at 90°.2 (m) (BMI = weight (kg)/height2 (m)).Weight and height were measured with an ASIMED MB201 scale and height rod with an accuracy of 100 g for weight (range 0.1–150 kg) and an accuracy of 0.5 cm for height (maximum capacity 200 cm). Weight was measured with the subject barefoot and wearing light clothing to the nearest 0.1 kg. Height was measured with the subject standing barefoot with heels together, arms at their sides, legs straight, shoulders relaxed and head aligned in the Frankfort horizontal plane with heels, buttocks, shoulder blades and the back of the head supported against the height rod. Measurements were recorded in centimetres. Body mass index was calculated as weight (kg) divided by heightWe conducted descriptive analysis of the variables of interest. We used means, standard deviations, medians and ranges to describe quantitative variables and the Kolmogorov–Smirnov test to check normality. We used proportions to describe qualitative variables and analysed linear association between the scores obtained in the short forms and the full version of the MNA test by Spearman’s test for correlations. Based on the categories “malnutrition”, “risk of malnutrition” and “normal nutritional status” we obtained Cohen’s kappa index to quantify agreement greater than that expected by chance between the MNA-SFs and full MNA classification. We obtained the confidence intervals for the estimated kappa values using the jackknife method which does not require normality for the variables . We perfBased on the categories “malnutrition and risk of malnutrition” vs. “normal nutrition status” and using the full MNA as the gold standard, we calculated sensitivity, specificity and predictive values in order to assess the diagnostic accuracy of the short forms of the MNA test. Additionally, we performed a non-parametric estimation of the ROC curve to evaluate the discriminatory power of the BMI-MNA-SF and CC-MNA-SF independently of the cut-off values. We used graphic representation and the Hanley and McNeil test to compare the diagnostic power of the two MNA-SFs. We performed statistical analyses with the SPSS  version 20.0 for Windows XP  and Epidat 4.2 .The proposal was reviewed by the Doctoral Committee of the Department of Nursing, Faculty of Health Sciences. The study has been approved by the Ethics Committee at the University of Alicante (File UA 2016-12-21).A formal written request to obtain permission for the study was made to the management teams at the community centres for the elderly included in it. All study subjects were asked to take part on a voluntary basis. They were properly informed about the study and their anonymity was guaranteed. Legal requirements and guidelines for good clinical practice and the World Medical Association Declaration of Helsinki on Ethical Principles for Medical Research involving Human Subjects (revised in October 2008) were met.The basic characteristics of the study population are shown in The mean scores we obtained in the three scales are high and are very similar in the two short scales. Out of a possible 14 points offered by both versions, the elderly respondents achieved an average score of 12.4 points (SD = 1.6) on the BMI-MNA-SF scale and 12.3 points (SD = 1.7) on the CC-MNA-SF scale. On the full version of the MNA, out of a possible 30 points they achieved a mean score of 25.3 points (SD = 2.4). The categorical classification given by the short versions shows that of the 479 persons classified with normal nutritional status by the BMI-MNA-SF and the 477 people classified with normal nutritional status by the CC-MNA-SF, in both cases there were 41 people at risk of malnutrition according to the full version of the MNA. In other words, had we not completed the full version of the scale, these 41 people  would have passed both screening tests without being considered at risk of malnutrition when in fact they were  and 2.ρ =0.78 p < 0.001; CC-MNA-SF: ρ = 0.78p < 0.001). There is also very high correlation between the two MNA-SFs .p < 0.001; CC-MNA-SF: κ = 0.52p < 0.001) indicate moderate agreement with the full MNA according to p < 0.001). If we compare the BMI-MNA-SF and CC-MNA-SF kappa coefficients using a kappa homogeneity test, we do not obtain any significant difference (p = 0.697).Since the point of the MNA is classifying free-living elderly in the three nutritional statuses , we completed this coefficient with a measure of agreement. The kappa coefficient indicates the agreement over and above the amount we would expect by chance. When interpreting the values of this coefficient according to the p < 0.001; CC-MNA-SF AUC =0.87p < 0.001) (p = 0.581). Considering the full MNA as the gold standard, the sensitivity obtained in the MNA-SFs is similar (73.4%) and the specificity is slightly higher in the BMI-MNA-SF than in the CC-MNA-SF (86.6% and 86.2%). Although prevalence does not affect sensitivity or specificity, the sensitivity value will decrease if the degree of malnutrition presented by the elderly is not very high. Given that the free-living elderly population resides at home, the prevalence of malnutrition is expected to be low when compared to hospitalised populations.In order to determine the ability of both MNA-SFs to identify subjects not requiring any nutritional intervention, we considered the dichotomised categorisation in the full MNA and the MNA-SFs as “risk of malnutrition” vs. “normal nutritional status”. Areas under the ROC curves using the full MNA as the gold standard reached high values (BMI-MNA-SF: AUC =0.88< 0.001) . We do nThe predictive values contained in In this study, we found that the BMI-MNA-SF and CC-MNA-SF scales have adequate predictive accuracy with respect to the MNA scale in non-institutionalized elderly. They showed high sensitivity and accuracy. We also obtained a moderate agreement between MNA and the two versions of the MNA-SF for identifying subjects with risk of malnutrition. In both cases, the overestimation of the nutritional status is only 6.2% compared to the MNA which is considered the gold standard. Malnutrition was not detected in the sample considered in our study  and the percentage of adults at risk of malnutrition was 23.3%.One limitation of our study is that the sample is not representative of the whole group of free-living elderly, as people with disabilities have been excluded.The strength of our study is that diagnostic accuracy tests of the CC-MNA-SF scale for an exclusively free-living elderly sample had not previously been conducted in Spain.Due to the high influence of the disease prevalence and its severity on statistical measures used to compare diagnostic tests , some stWe have performed an exhaustive revision of recently studies performed on the MNA, BMI-MNA-SF or CC-MNA-SF in non-institutionalized elderly populations or in mixed populations. At the international level, we have found one study performed in Germany and Italy with 675 elderly people divided into three groups: community-dwelling older people, nursing homes residents and patients in geriatric rehabilitation , concernIn Spain we found two studies  where thThe only study we have found in Spain, where the CC-MNA-SF is compared to the MNA test is the one by It may be observed in former studies, that when we are working on a population where de prevalence of malnutrition is low, as in our case, the existence of an imbalance between the percentage of elderly people in a normal nutritional state and the percentage of elderly people at risk or malnourished, does the random agreement be greater and the kappa coefficient has smaller values. Also, the low prevalence of malnutrition has negative influence on the predictive positive value and positive on the negative predictive value. On the other hand, even though prevalence does not have a direct influence on the sensitivity and specificity of diagnostic test, these values are influenced by the level of malnutrition, because of the difficulty of detecting elderly people with mild level of malnutrition.This justifies the importance of doing studies on these diagnostic tests on different populations to be able to get proper decisions based on the results obtained. On the few studies done up to this moment, in populations with a malnutrition prevalence and a level of malnutrition similar to that of the patients analysed on this study, results are similar to ours. Also, we think this is one of the first studies where the CC-MNA-SF test has been analysed with non-institutionalized elders in Spain.Bearing in mind that the free-living elderly population is growing and that inexpensive, quick and efficient nutritional assessment tools are needed for use in primary healthcare settings, we suggest that the CC-MNA-SF scale is a useful screening tool and a valid alternative to the BMI-MNA-SF, although the characteristics of the population must always be taken into account to make the right decisions based on the MNA-SF scales. Since malnutrition prevalence is usually not very high in this type of population, the level of agreement between the BMI-MNA-SF, CC-MNA-SF and the MNA will be moderate. On the other hand, given that malnutrition is not usually present at an advanced stage in free-living elderly, the sensitivity of short-forms scales also may be affected. Therefore, when the CC-MNA-SF results are not in accordance with other additional information and the doctor is in doubt, the proper decision will be to complete the full MNA or perform additional examinations.To sum up, in our study, both versions of the MNA-SF have high predictive ability to identify nutritional risk in free-living elderly people. In both cases there was an overestimation of their nutritional status of only 6.2%. The new version (CC-MNA-SF) is easier to use as the BMI parameter can be replaced by CC in primary care centres, taking less time and being less costly, as it does not require scales and a large equipment to measure weight and height.However, more studies would be needed in order to better understand their discriminatory capacity in non-institutionalised elderly populations. It would also be useful to study the problems to identify potential malnutrition in this type of population, since its degree of malnutrition is low and also to investigate the possibility of raising the CC-MNA-SF cut-off point for this group.10.7717/peerj.3345/supp-1Supplemental Information 1Click here for additional data file.


Diagnosticity proponents have offered several criticisms of ROCs, which we show are either false or irrelevant to the assessment of eyewitness accuracy. We also show that, like diagnosticity, Bayesian measures of identification accuracy confound response bias with witnesses’ ability to discriminate guilty from innocent suspects. ROCs are an essential tool for distinguishing memory-based processes from decisional aspects of a response; simulations of different possible identification tasks and response strategies show that they offer important constraints on theory development.How should the accuracy of eyewitness identification decisions be measured, so that best practices for identification can be determined? This fundamental question is under intense debate. One side advocates for continued use of a traditional measure of identification accuracy, known as the  The accuracy of the identification itself has almost always been summarized with a measure known as diagnosticity, which is simply a ratio of correct and incorrect identification probabilities. Crucially, recent research has demonstrated that diagnosticity is confounded with witnesses’ willingness to identify a suspect, raising questions about the validity of prior conclusions about the effects of system and estimator variables. Those conclusions rest on the assumption that manipulations of these variables affect only the accuracy of the identification, but the diagnosticity measure cannot distinguish between an effect on accuracy and an effect on identification willingness to identify. More appropriate measures of eyewitness identification accuracy can be obtained from receiver operating characteristic curves (ROCs) based on confidence ratings. Although ROCs have a long and productive history in many domains of psychological research (including recognition memory), their application to eyewitness identification decisions has been challenged. In this paper, we correct some of the conceptual errors in the literature on identification ROCs, and demonstrate their value for both understanding the factors that influence eyewitness identification accuracy and constraining theoretical development.Eyewitness identifications can provide compelling evidence in criminal cases. For decades, researchers have explored the factors that influence the accuracy of those memory judgments, including consideration of both H/F; Wells & Lindsay, diagnosticity ratio, or simply as diagnosticity. The other side of the argument is that diagnosticity is confounded with witnesses’ tendency to identify someone; that is, to “choose” from the lineup  against the probability of a false identification of an innocent suspect (F) at each confidence level or response bias. By definition, the empirical points on any given ROC differ only in terms of response bias, whereas ROC that fall higher in the space reflect greater ability to discriminate guilty from innocent suspects because the hit rate is higher for any particular false alarm rate. Levi  lineup (including filler identifications), as Wells and colleagues proposed  , contribute to determining the optimal decision criterion or response bias  all implicate the innocent suspect, whereas in a fair lineup procedure, those errors are distributed across the suspect and the fillers  is reflected at every confidence level.1. In contrast to Lampinen’s  was equated for two different identification tasks: showups and six-photograph lineup identifications. In both cases, Lampinen assumed an underlying representation based on signal detection theory Lampinen  argued tTo simulate the showup procedure, Lampinen randomly selected a single value from either the guilty suspect distribution, to represent a TP showup, or the foil distribution, to represent a TA showup. The sampled strength was compared with a fixed set of criterion values  to assign a confidence level for the response. For example, a sampled strength of 1.0 would be assigned a confidence level of 6 because it falls in the interval between the fifth and sixth criterion locations. Any sampled strength greater than 0.18, the lowest criterion, was assumed to result in a positive identification.Simulation of the six-photograph lineup procedure was similar, except that each simulated lineup involved six sampled strengths, either one from the guilty suspect distribution and five from the filler distribution (for a TP lineup) or all six from the filler distribution (for a TA lineup). In either case, the highest sampled strength determined the confidence rating. If that strength was from the guilty suspect distribution, then a “hit” resulted. If the lineup included only fillers, then any strength above the lowest criterion (0.18) was treated as a false alarm. Because there are six opportunities for a filler to exceed the criterion in a TA lineup, the resulting response rates were divided by 6.d′ values he considered. Because the AUC is a measure of subjects’ ability to discriminate between two types of stimuli  actually varies.Lampinen  showed tWe performed simulations similar to Lampinen’s so that the AUCs could be calculated. To extend and generalize his analyses, we varied the criterion locations continuously for 10,000 different simulated trials of each type, thus mapping out the full theoretical ROC. By comparison, Lampinen assumed a particular set of fixed criterion locations, which necessarily provides only a snapshot of the ROC. In addition, we assumed several different lineup sizes . Although lineups with only two or three photographs are not used in standard police procedures, their inclusion in our simulations does provide some insight into the relationship between lineup size and form of the ROC.d′ of 1.5 and equal variance distributions.  Notice that these four ROCs are visually indistinguishable at the lowest false alarm rates (up to about 0.10) that reflect the highest confidence levels. This part of the ROC was not shown in any of Lampinen’s simulations, because the most conservative criterion he selected was relatively liberally placed and thus produced high hit and false alarm rates overall. The operating points in Lampinen’s simulations are marked with red open circles on the showup ROC in Fig. Our simulated showup and lineup ROCs are shown as the set of ROCs in Fig. d′) is equated. Lampinen’s selection of particular, relatively liberal decision criteria obscured this similarity.The highest-confidence identification decisions are those that are most valuable to the legal system because they carry the greatest probative value , the greater the number of fillers in a TP lineup, the more likely it is that one of them will have greater familiarity than the guilty suspect just by chance. In that case, the witness would choose a filler instead of the guilty suspect, thus reducing the hit rate. A consequence of these two effects is that the very same evidence values in memory, used in conjunction with identical criterion locations, can nonetheless result in hit and false alarm rates that are reduced as the number of photographs in the lineup increases. For example, the blue circles in Fig. d′ value of 1.5 and use a decision criterion of 1.00 for both showup and six-person lineup decisions: Their decisions appear more conservative in the lineup task simply as a consequence of the number of photographs presented .At lower levels of confidence, or more liberal response biases, it becomes apparent that the number of response options affects both the hit and false alarm rates. The false alarm rate is limited for the obvious reason that selection of any of the fillers in a TA lineup is an error; chance response rates are properly limited to 1/d′), selecting the pAUCs at the 2.5 and 97.5 percentiles as the lower and upper bounds of the 95 % confidence intervals.To estimate the area under the curve for each of the ROCs shown in Fig. d′) is equated.smaller gray square in Fig. d′) varies, the pAUC for both showup and six-person lineup identifications capture those changes quite readily. Figure d′ units, pAUC changes are obvious for both showup and lineup decisions  is held constant and change appropriately (and quite similarly) when true accuracy varies.Figure assume witnesses are equally good at discriminating guilty from innocent suspects in showup and lineup tasks. On one hand, they show that, theoretically, if discrimination accuracy (d′) is the same, the ROCs are highly unlikely to suggest that either of the procedures results in higher-accuracy decisions than the other. On the other hand, the simulations also show that real differences in discrimination accuracy (d′) can be detected with ROCs, and to the same degree for both showups and lineups. Whether accuracy is actually the same in these two identification procedures is an empirical question, not a theoretical one. Empirical comparisons of ROCs for showup and lineup identifications have consistently revealed that showups result in significantly lower decision accuracy than lineup procedures, as measured with pAUC  affected estimated but not true witness accuracy, but the primary application of ROCs to eyewitness identification decisions has been to compare sequential and simultaneous presentation of the same lineup photographs. The consistent finding, that pAUC for simultaneous lineups is equal to or greater than for sequential procedures , where s is the standard deviation of the guilty suspect distribution.d′ of 1.5 and s = 1, is shown as the solid curve in Fig. Finally, note that the “proof” offered by Lampinen ,N,d′) is equated. Thus, Lampinen’s argument does not change either the empirical conclusion that simultaneous lineups yield equal or greater AUCs than sequential lineups  differs systematically for showups and lineups. Our analyses demonstrate that the two paradigms yield essentially identical estimates of performance when the true accuracy level (d′) would be observed for lineup and showup identifications. Finally, Lampinen notes that if response bias differs across identification procedures, then the operating points being compared may reflect different degrees of witness confidence. Conversely, Levi (d′).Lampinen’s  second mly, Levi  worries same accuracy (d′) at every point. Because of this property of ROCs, one can readily see both accuracy and response bias effects simultaneously: Curves higher in the space reflect higher (d′) decision accuracy , and points toward the lower-left end of the curve reflect more conservative responses. One does not need to compute a single-point measure of accuracy, such as d′, at a given false alarm rate to compare accuracy across two conditions. Indeed, one should not do so, because d′ is confounded with response bias whenever the underlying evidence distributions have unequal variance, as is consistently observed in recognition memory judgments  conclude that accuracy differs if they select an inappropriate accuracy measure . For this reason, diagnosticity is not an appropriate measure of eyewitness identification accuracy  that yield the high-confidence responses that are most useful to the legal system  that is independent of response bias. In contrast, single-point measures, such as information gain they reported is the difference in the probability that the suspect is actually guilty before and after an identification (or lineup rejection) occurs . The information gain curves they reported vary over the probability that the lineup includes a guilty suspect  and as a function of witness confidence.Bayesian measures of eyewitness identification decisions have also been proposed  in each task. For example, Wells et al. (d′) change as a function of filler similarity. The posterior probability of guilt and information gain plots do not allow us to distinguish between these two interpretations of the data: Both discrimination and response bias contribute to each curve. Similarly, Wells et al. , each with two different response biases or confidence levels that were selected for illustrative purposes. We then used those data to calculate the posterior probability of guilt and the information gain curve for each combination of discrimination and bias; these are displayed in Fig. if the response bias is conservative. This effect may explain the information gain curves derived from sequential and simultaneous lineups , the witness’s willingness to identify someone (response bias or confidence), and the true discrimination potential provided by the identification procedure. As we have seen, ROCs provide the best way of independently assessing response bias and witness discrimination of guilty from innocent suspects.In this final section, we suggest some directions for the field in which ROCs might play a particularly important role. As Lampinen  also notTo demonstrate some of the potential of ROCs for both constraining theory development and guiding decisions about experimental paths to pursue or to avoid, we simulated a few different decision strategies that have been proposed in the literature. The simulated data were then used to generate the ROC that would result if that hypothetical decision process were used; in principle, these theoretical ROCs can be compared with empirical data. This approach can be particularly useful for eliminating potential decisional processes from further consideration if, for example, the theoretical ROC fails to correspond to empirically observed ROCs.In our first simulation, we considered two decision rules that have been proposed for identification decisions in a simultaneous lineup procedure, namely the absolute strength rule, in which the strongest above-criterion photograph is selected Clark, , and a rd′ = 1.5 and a standard deviation of 1, and five values were sampled from a distribution with a mean of 0 and standard deviation of 1; for TA lineups, six values were randomly sampled from an N distribution. This sampling process was repeated 10,000 times. The criterion location was varied in small steps over a wide range; for the relative decision rules, the minimum difference in strengths was arbitrarily set to 0.15, one-tenth the true discrimination value. The results, shown in Fig. For these simulations, we used the same sampling process as we described previously. For TP lineups, one value was sampled from a normal distribution with a mean of d′ = 1.5.In our second simulation, we compared a six-person simultaneous lineup task with a sequential identification procedure using the same sampled evidence values. Simulation of the simultaneous task was identical to that used to generate Fig. red; in this region, the criterion location is placed no more conservatively than d′/2.) In the limit, if the criterion is set so low that any photo is given a positive response, then the hit rate must simply equal the probability that the guilty suspect appears first in the sequence, or one in six for our simulated six-person lineups. Importantly, the same memory strengths were used on both simulated lineup tasks, and thus true discrimination of guilty from innocent suspects is equated for these ROCs. The curves diverge only because of differences in the specific decision processes we simulated.Although the simulated simultaneous lineup ROC looks quite similar to those observed empirically  and does not provide information that is redundant with diagnosticity. ROCs do not “compare hit rates after equating false alarm rates”  at all possible response biases. This clear advantage of ROC analyses has fueled theoretical advances in other domains, including recognition memory (Dougal & Rotello, ROCs


Previous research documents that men and women can accurately judge male physical strength from gait, but also that the sexes differ in attractiveness judgments of strong and weak male walkers. Women’s (but not men’s) attractiveness assessments of strong male walkers are higher than for weak male walkers. Here, we extend this research to assessments of strong and weak male walkers in Chile, Germany, and Russia. Men and women judged videos of virtual characters, animated with the walk movements of motion-captured men, on strength and attractiveness. In two countries (Germany and Russia), these videos were additionally presented at 70% (slower) and 130% (faster) of their original speed. Stronger walkers were judged to be stronger and more attractive than weak walkers, and this effect was independent of country (but not sex). Women tended to provide higher attractiveness judgments to strong walkers, and men tended to provide higher attractiveness judgments to weak walkers. In addition, German and Russian participants rated strong walkers most attractive at slow and fast speed. Thus, across countries men and women can assess male strength from gait, although they tended to differ in attractiveness assessments of strong and weak male walkers. Attractiveness assessments of male gait may be influenced by society-specific emphasis on male physical strength. Physical strength is a sexually dimorphic, heritable trait  with menPrevious research documents that men and women can accurately assess male physical strength from men’s faces, bodies, and voices , 2010, a1). There is little information available about how gait is perceived in different countries, and most investigations of gait are clinical or sports science studies. In a preliminary study on Brazilian and German women’s attractiveness perceptions of (British) men’s dance movements, In the present study, we expand on these findings by considering men’s and women’s perceptions of male gait in Chile, Germany, and Russia – three countries that differ in social and cultural norms (among other measures), and along indices of development  as part of a large-scale study on body movement in relation to anthropometry and personality  running Vicon Nexus software. Thirty-nine reflective markers were attached to each participant’s major joints and body parts (Plug-in-gait marker set). Participants did not receive instructions on how to walk, but were told to remain within a certain area (marked with adhesive tape on the floor) in a room dedicated to motion-capturing. A male and a female investigator was present during recordings. Gait recordings were applied to size- and shape-standardized, sex-neutral humanoid characters using Motionbuilder software  and rendered as 773 pixel × 632 pixel video clips. A sequence of 3 s (4–5 strides) was digitally isolated from the middle of each walk sequence.M = 21.6, SD = 4.1), the videos of the 10 strongest  and 10 weakest participants   were selected for the rating study. Strong and weak walkers differed in HGS but not in the number of strides displayed in the videos .Participants handgrip strength  was measured with a hand dynamometer , twice for each hand, and the means of the two left and two right HGS measurements were used for assignment to either the “strong” or “weak” walker group. For the selection of strong and weak walkers, we considered only participants for which no issues in post-processing of walk movements were noted . Of the remaining 70 men  judged slow, normal, and fast versions of the videos for “naturalness” on a 4-point Likert-type scale . Naturalness judgments did not differ across conditions .In addition to the videos displaying strong and weak male walkers at “normal” speed, two further sets of these walkers were created by manipulating playback speed, thus rendering the gait “slower” and “faster.” Slow and fast versions were created by setting playback speed to 70% (slow) and 130% (fast) of the original video using iMovie software . To test for artifacts in manipulations, 15 German men and women  were used to construct a new video showing walk movements in a loop.n = 122; 60 males, M = 23.0, SD = 3.4; 62 females, M = 24.4, SD = 3.7), the Universidad de Playa Ancha  and the Russian State University for the Humanities .The gait videos were shown to 188 men and 199 women (aged 16 to 50 years), recruited at the University of Göttingen . Gait assessments at normal speed were collected in Chile, Germany, and Russia. The total sample size for country comparisons of strong and weak walkers at normal speed was n = 213. Additional assessments of strength and attractiveness for walking speed manipulations were collected in Germany  and Russia . The total sample size for country comparisons of strong and weak walkers varying in gait speed was n = 302.Participants judged the walk movements for strength and attractiveness using a 7-point scale (1 = 2. The videos were embedded in the presentation via a (private) YouTube account with advertisements removed and could be replayed. Judgments of attributes were made in blocks and the order of clips was randomized within each block, with blocks randomized across participants.In Germany, video ratings were collected on laptop computers running Medialab v2012  software. In Chile and Russia, videos were presented on PC monitors using SurveyMonkeyFigure 1 presents descriptive statistics for men’s and women’s strength and attractiveness perceptions of strong and weak male walkers in Chile, Germany, and Russia. A mixed-design ANOVA was performed on perception measures with walker strength as a within-subjects factor and country and observer’s sex as between-subjects factors (Table 1). For this analysis, we considered only judgments of walkers in the “normal speed” condition. There were main effects of walker strength on perceptions of strength and attractiveness. Strong walkers were rated higher on both attributes . The interaction effect of walker strength × country was significant for perceptions of attractiveness but not strength. Gait attractiveness ratings of strong walkers were highest in Germany, followed by Russia and Chile; for weak walkers, Russian attractiveness ratings were highest, followed by German and Chilean assessments . However, post hoc multiple tests did not reveal significant differences for these comparisons .Figure 1), although these differences were not statistically significant . Both sexes gave higher strength ratings to strong walkers as compared to weak walkers . There was a three-way interaction effect of walker strength × country × sex for perceptions of attractiveness but not strength (Table 1). In Chile and Germany, women’s attractiveness judgments of strong male walkers were higher than those of men whereas men judged weak walkers more positively than did women. In Russia, men gave lower attractiveness scores to weak walkers as compared to women . However, paired t-tests did not reveal significant differences for these comparisons .An interaction effect of walker strength × sex was detected for perceptions of attractiveness but not for strength. Women judged strong walkers higher on attractiveness than did men, and men judged weak walkers higher on attractiveness than did women . There were no interaction effects of walker strength × walking speed on perceptions of strength and attractiveness. However, the interaction effect of walker strength × walking speed × country was significant for perceptions of attractiveness but not strength. German and Russian participants rated strong walkers higher on attractiveness than weak walkers when viewed at slow and fast speed. At normal speed, German attractiveness ratings of strong walkers were higher than those of Russians, and for weak walkers, Russian attractiveness ratings were higher than those of Germans. However, post hoc multiple tests did not reveal significant differences for these comparisons  except for German attractiveness perceptions of weak walkers at normal vs. slow speed . No significant 4-way interaction effect of walker strength × walking speed × country × sex was detected for perceptions of strength and attractiveness.The results of the present study confirm the previously reported effect of male physical strength on perceptions of strength and attractiveness from gait  in threePerceptions of strength distinguished between strong and weak male walkers in all three countries. Perceptions of attractiveness varied between countries and sex, although effect sizes for these interactions were small. Thus, interpretations of country- and sex-specific assessments of male gait attractiveness must be made with caution. In Chile and Germany, women judge strong male walkers to be more attractive than weak male walkers, whereas in Russia there was no such difference. In the current study, we focused on comparisons of men’s and women’s perceptions of strength from gait across countries. Thus, we did not include specific measures of, for example, social dominance orientation SDO;  and otheManipulations of playback speed of gait did not affect strength perceptions of men or women in either country in which they were investigated (Germany and Russia). Thus, the assessment of physical strength from gait cues appears to be independent of gait speed. German and Russian participants judged strong walkers as more attractive when viewed at slow and fast speed. Germans tended to judge strong walkers higher on attractiveness than Russians when viewed at normal speed and Russian attractiveness judgments of weak walkers were higher than those of Germans. Again, because of the small effect sizes, these results should be interpreted with caution. The manipulation of playback speed of gait videos should be regarded as approximation to “ideal” stimuli, thus providing only preliminary information about perceptions of gait speed. Walking faster or slower is associated with changes in movement biomechanics, and such changes were not considered in our manipulations. Future research may profitably investigate whether it is, in fact, gait speed changes that influence women’s attractiveness perceptions of male walkers (as suggested by the current study), or if altering gait speed is linked to other gait characteristics that affect perception.We demonstrate an effect of male gait on perceived strength in three countries, suggesting that the capacity to derive physical strength information from male gait is not culture-specific. Women more than men seem to discriminate between strong and weak walkers when assessing attractiveness of male gait. Attractiveness assessments of male gait show some local variation, and may be influenced by society-specific emphasis of male physical strength. Future studies investigating the effect of strength from gait cues on perception should test whether the reported effects extend to pre-industrialized societies. If this is the case, it would support the suggestion that specific biomechanical characteristics of male gait are linked to physical strength, and that humans may be adapted to use these cues in social perception.This study was carried out in accordance with the recommendations of the German Psychological Society. All subjects gave written informed consent in accordance with the Declaration of Helsinki. In case of minors, consent was additionally obtained from their legal guardians or teachers. The protocol was approved by the ethics committees of the University of Göttingen and Northumbria University.BF, MW, and JO conceived the study. MW, MB, AM, and JM-R collected data. BF and MW analyzed the data. BF, JO, MB, YS, and TS wrote the manuscript. All authors edited the manuscript for intellectual content, provided critical comments on the manuscript, and agreed to be accountable for the content of the work.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Recent studies have shown that mathematical models can be used to analyze brain networks by quantifying how likely they are to generate seizures. In particular, we have introduced the quantity termed brain network ictogenicity (BNI), which was demonstrated to have the capability of differentiating between functional connectivity (FC) of healthy individuals and those with epilepsy. Furthermore, BNI has also been used to quantify and predict the outcome of epilepsy surgery based on FC extracted from pre-operative ictal intracranial electroencephalography (iEEG). This modeling framework is based on the assumption that the inferred FC provides an appropriate representation of an ictogenic network, i.e., a brain network responsible for the generation of seizures. However, FC networks have been shown to change their topology depending on the state of the brain. For example, topologies during seizure are different to those pre- and post-seizure. We therefore sought to understand how these changes affect BNI. We studied peri-ictal iEEG recordings from a cohort of 16 epilepsy patients who underwent surgery and found that, on average, ictal FC yield higher BNI relative to pre- and post-ictal FC. However, elevated ictal BNI was not observed in every individual, rather it was typically observed in those who had good post-operative seizure control. We therefore hypothesize that elevated ictal BNI is indicative of an ictogenic network being appropriately represented in the FC. We evidence this by demonstrating superior model predictions for post-operative seizure control in patients with elevated ictal BNI. In practice, BNI can be calculated as the average proportion of time that network nodes spent in the seizure-like state during a sufficient long simulation time. BNI has been shown to be a useful biomarker of idiopathic generalized epilepsy . The surgery outcome was measured in terms of seizure rate using the Engel scale . The signals were recorded intracranially by grid, strip, and depth AdTech electrodes  and a NicoletOneTM recording system . This study was approved by the Internal Review Board of the Inselspital . All patients gave written informed consent that imaging and EEG data may be used for research purposes. Other details about this dataset have been previously described .U-test to assess whether the correlations in the original time series (ρ0) were significantly different compared with the correlations in the surrogates (ρsurr), and we applied Bonferroni–Holm corrections to account for multiple comparisons. We thus obtained a surrogate-corrected correlation matrix using the formula:s = 1 if the null hypothesis of the statistical test was rejected, or s = 0 otherwise . The iEEG signals were down-sampled to 512 Hz and re-referenced against the median of all artifact-free segments as judged by visual inspection (Kaspar Schindler). These data were then band-pass filtered between 0.5 and 120 Hz and notch-filtered between 48 and 52 Hz. Each epoch was divided into a set of 8 s segments, with 1 s gaps between each of them. The gaps were pragmatically chosen to be 1 s to provide a balance between the need for sufficient sampling and computational tractability. Next, we computed 10 univariate iterated amplitude adjusted Fourier transform surrogates for each segment. The segments were then divided in 10 subsegments of 2 s distributed with minimal overlap. This procedure resulted in an ensemble of 10 subsegments for the original time series, and 100 subsegments for the surrogates. The FC was inferred from the correlations between the time series of each iEEG channel using the Pearson’s equal-time (zero-lag) cross-correlation coefficient ρ. We used the Mann–Whitney–Wilcoxon therwise , 24. Theh2 index , i.e., uj and it obeys the ODE:jI(t) accounts for incoming currents,j denotes node j , K is the global scaling coupling, ija is the i,jth entry of the adjacency matrix (the FC network), and i  and evaluating the time spent in spiking dynamics. Spikes were extracted by applying a threshold to the phases as described in Ref.  and BNI upon removal of the sn nodes. In the model, a successful surgery will have higher ΔBNIns than an unsuccessful one. The set sn of nodes removed were identified from the coregistration of high resolution T1-weighted MR images acquired before and after surgery in all individuals [see the full details of these procedures in Ref.  of the receiver operating characteristic. Having obtained two AUC, we wanted to compare them. We used all possible combinations of half-group splitting of ΔnsBNI to find a distribution of AUC values . We then compared the two distributions using the Mann–Whitney–Wilcoxon U-test.To test for statistically significant differences between ictal BNI and pre- and post-ictal BNI of the whole cohort of patients, we used the Kruskal–Wallis test. Differences were considered statistically significant when the p < 0.001, Kruskal–Wallis test). Equivalent results were also observed for FC inferred using the non-linear h2 index . Specifically, we observed a sharp increase in BNI at seizure onset, but, interestingly, BNI remained at similar levels to those of the ictal epoch for about 1 min into the clinically defined post-ictal epoch. These results were robust for different choices of the strength of coupling between nodes in the BNI calculation .Figure U-test was performed to assess whether the median BNI in the ictal period was larger than the pre-ictal median BNI. We found elevated ictal BNI in both peri-ictal epochs of eight individuals, with the remaining eight individuals not displaying a significant increase in BNI during the ictal period in at least one of the peri-ictal epochs. Figure S3 in Supplementary Material shows the BNI evolution of four representative individuals, two exhibiting consistent elevated ictal BNI in both peri-ictal epochs, and two presenting inconsistent peri-ictal BNI. We further found that elevated ictal BNI was correlated with post-operative outcome . In five of six people who were seizure free post-surgery (Engel I), we observed a significant increase in BNI in ictal epochs relative to the pre-ictal epochs .We next examined the pattern of BNI evolution for each of the two peri-ictal epochs of each individual separately. A one-tailed Mann–Whitney–Wilcoxon see Ref.  for moreTo date, the concept of BNI—that is the time spent in the seizure state averaged across each node within a network—has proved to be valuable for both epilepsy diagnosis –19 and ah2 index . Future work will also therefore investigate the ways in which linear and non-linear channel associations evolve at seizure termination.It has previously been shown that FC topology changes during seizures –24. HereThe emerging field of network model-based recommendations for epilepsy surgery is based on a number of assumptions. First, it assumes the existence of an ictogenic network, i.e., a brain network responsible for the generation of seizures. In fact, there is extensive neuroimaging and iEEG evidence for the widespread involvement of brain regions in focal epilepsies –6, 27. HThis study was carried out in accordance with the recommendations of the Internal Review Board of the Inselspital with written informed consent from all subjects. All subjects gave written informed consent in accordance with the Declaration of Helsinki. The protocol was approved by the Internal Review Board of the Inselspital .ML, MG, and JT: study concept and design, results interpretation, and manuscript drafting and revision. ML: data analysis. MR: results interpretation and manuscript revision. EA, CR, and KS: acquisition of clinical data, results interpretation, and manuscript revision.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Mycobacterium marinum infection in adult zebrafish resembles human TB. Here, we present a pre-clinical screen for a DNA-based tuberculosis vaccine in the adult zebrafish using an M. marinum infection model. We tested 15 antigens representing different types of mycobacterial proteins, including the Resuscitation Promoting factors (Rpf), PE/PPE protein family members, other membrane proteins and metabolic enzymes. The antigens were expressed as GFP fusion proteins, facilitating the validation of their expression in vivo. The efficiency of the antigens was tested against a low-dose intraperitoneal M. marinum infection (≈ 40 colony forming units), which mimics a primary M. tuberculosis infection. While none of the antigens was able to completely prevent a mycobacterial infection, four of them, namely RpfE, PE5_1, PE31 and cdh, led to significantly reduced bacterial burdens at four weeks post infection. Immunization with RpfE also improved the survival of the fish against a high-dose intraperitoneal injection with M. marinum (≈ 10.000 colony forming units), resembling the disseminated form of the disease. This study shows that the M. marinum infection model in adult zebrafish is suitable for the pre-clinical screening of tuberculosis vaccines and presents RpfE as a potential antigen candidate for further studies.Tuberculosis (TB) remains a major global health challenge and the development of a better vaccine takes center stage in fighting the disease. For this purpose, animal models that are capable of replicating the course of the disease and are suitable for the early-stage screening of vaccine candidates are needed. A  Mycobacterium tuberculosis, the causative agent of TB, caused 1.4 million deaths and 10.4 million new infections in 2015 [M. tuberculosis strains and HIV co-infections hamper the treatment of TB [Tuberculosis (TB) remains a major health problem that has been extensively studied in recent years. Interest in developing novel tuberculosis vaccines has grown over the years. The only available TB vaccine, Bacillus Calmette Guérin (BCG), protects young children, but its ability to induce long-term cell mediated immune responses varies and the protection it provides against pulmonary TB or against the reactivation of latent TB is limited –5. ThereM. tuberculosis is not a natural pathogen of traditional animal models such as mice, rabbits and guinea pigs, and natural hosts, non-human primates, can be used only very selectively for experiments [Danio rerio) has emerged as an advantageous animal to model a TB infection. An infection with Mycobacterium marinum—a close relative to M. tuberculosis—in zebrafish leads to a disease that resembles human TB in many aspects [M. marinum is a natural pathogen of fish and an infection can lead to either an active or a naturally latent form of the disease [M. marinum and M. tuberculosis [A central issue in TB research has been the paucity of good animal models . M. tubeeriments . In the  aspects . M. mari disease –10; revi disease . As a ve disease –17. Alth disease . In addirculosis ,27–31. Trculosis ,25,30,32Despite the increasing knowledge on mycobacterial pathogenesis, the development of new TB vaccines has turned out to be challenging. Currently there are 14 vaccine candidates in the clinical trial pipeline, including inactivated or attenuated whole-cell vaccines, and subunit vaccines containing mycobacterial antigens. . Expressin vivo infection models are required to assess the efficacy of the novel vaccine candidates as there are currently no reliable biomarkers for predicting the efficacy of protection against TB [M. marinum infection model in the pre-clinical screening of DNA-based tuberculosis vaccines. Based on literature and online databases, MarinoList and TubercuList [M. marinum antigens that have a homologue in M. tuberculosis and predicted or experimentally shown immunogenicity. We selected molecules that belong to different functional categories and are expressed during different stages of mycobacterial growth, including four Resuscitation promoting factors [A key step in the design of DNA vaccines is the choice of the antigen(s), especially since DNA vaccines tend to have a relatively weak immunogenicity in humans . Even thainst TB ,36,37. Wainst TB ,39. The ercuList ,41, we s factors , three P factors  and fiveAdult (5–7 month-old) wild type AB zebrafish were used for all experiments and maintained as in . Animal studies were approved by the National Animal Experiment Board in Finland  and conducted in accordance with the EU-directive 2010/63/EU on the protection of animals used for scientific purposes.Mycobacterium marinum strain ATCC 927 was cultured on 7H10 Middlebrook OACD plates  at +29°C, inoculated to a fresh plate every 7 days, and a fresh stock was thawed after every two passages. Liquid cultures for RNA isolation and infections were grown in 7H9 Middlebrook medium , see below for details.The M. marinum genes corresponding to the selected antigens was confirmed by qRT-PCR from M. marinum RNA. The M. marinum ATCC 927 strain was cultured in 7H9 (BD Biosciences) medium to the log phase (OD600 of 0.6). Bacteria from six separate liquid cultures were collected by centrifuging for 5 minutes at 800 x g. The pellets were used for RNA extractions with the RNeasy Mini Kit (Qiagen) according to the manufacturer’s instructions. Before qRT-PCR, the RNA samples were treated with DNase . The expression of mycobacterial genes was verified with the iScript™ One-Step RT-PCR Kit with SYBR® Green  according to the manufacturer’s instructions. The M. marinum internal transcribed spacer (MMITS) [Expression of the  (MMITS) . The pri (MMITS)  and are M. tuberculosis and M. marinum genes was analyzed with the Clustal Omega sequence alignment tool [M. marinum proteins was determined based on experimental evidence available in the literature or by prediction of transmembrane protein topology with a hidden Markov model [M. marinum protein was included in the vaccine antigen. The Expasy Compute pI/Mw tool [Homology between the ent tool . The celov model . Antigen/Mw tool  was usedMycobacterium marinum ATCC grown on 7H10 Middlebrook OACD plates  by colony PCR. Purified PCR products were restriction cloned into the pCMV-eGFP expression vector to be expressed with a C-terminal GFP tag (Addgene plasmid 11153), transformed into E. coli One Shot TOP10 cells (Invitrogen) and confirmed by sequencing. For DNA immunizations, plasmid DNA was purified using the QIAGEN Plasmid Plus Maxi Kit . The pCMV-EGFP plasmid without mycobacterial inserts was used for control vaccinations.DNA vaccine constructs were prepared and the DNA vaccinations performed as described in . In briepCMV-eGFP plasmid using aluminosilicate capillary needles and a PV830 Pneumatic PicoPump microinjector . The injection was followed by electroporation  using the GenePulser-electroporator  with tweezer-type electrodes  [For vaccine immunizations, the fish were briefly anaesthetized in 0.02% 3-aminobenzoic acid ethyl ester (pH 7.0) (Sigma–Aldrich) and injected in the dorsal muscle with 12 μg of the vaccine or the ton, MA) .In vivo expression of the plasmid DNA-derived protein products (GFP and its antigen recombinants) was verified by fluorescence microscopy, Western blotting and ELISA using naïve fish as a negative control. Nikon AZ100 fluorescent microscope was used for the microscopy. For Western blotting and ELISA, the fish were dissected under UV light and their dorsal muscles that showed the fluorescence indicative of vaccine antigen expression were collected for analysis. The samples were homogenized in TriReagent  with ceramic beads  using a PowerLyzer™ 24 Bench Top Bead-Based Homogenizer (MO BIO Laboratories), followed by a protein extraction protocol according to the manufacturer’s instructions.® BCA Protein Assay Kit  was used to define the total protein concentration of each lysate. For Western blotting, a volume corresponding to a total protein amount of 7.5–15 μg of each fish homogenate was resolved on a 4–20% Mini-PROTEAN® TGX™ Gel (BioRad) and blotted onto a nitrocellulose membrane using Trans-Blot® Turbo™ Mini Nitrocellulose Transfer Packs (BioRad). The horse radish peroxidase conjugated GFP Tag Monoclonal Antibody (GF28R) (Thermo Fisher) was used for detection of the target protein. The GFP ELISA Kit  was used for determining the relative levels of GFP according to the manufacturer’s instructions. The absorbance values were transformed into GFP concentrations using a GFP standard, and the amount of GFP in each sample was normalized with the total protein concentration of the sample and with the average of the GFP controls in the experiment. Non-immunized AB fish were used as a negative control in both Western blots and ELISA.The PierceM. marinum four weeks after immunization. M. marinum ATCC 927 was cultured at 29°C in standard mycobacterium medium, 7H9 (BD Biosciences), and prepared for infections as described in [M. marinum diluted in 0.2 M sterile KCl was injected intraperitoneally (i.p.). Thereafter, the fish were immediately released into a recovery tank. Infection doses were verified by plating the bacteria onto 7H10 plates (BD Biosciences). Following the infections, the well-being of the fish was monitored daily, and fish showing signs of stress or mycobacterial disease during the experiment follow-up period were euthanized with 0.04% 3-aminobenzoic acid ethyl ester.Fish were infected either with a low (~40 cfu) or high  dose of ribed in . For infM. marinum four weeks post-immunization. Five weeks after infection, the fish were subjected to DNA extraction. Fish showing signs of disease during the five-week follow-up interval were euthanized immediately and included in the cfu analysis. The contents of the peritoneal cavity, including the visceral organs, of euthanized fish was collected into homogenization tubes  and homogenized in 1.5 ml of TRI reagent  using the PowerLyzer24 bead beater (Mobio). Homogenized samples were sonicated using an m08 water bath sonicator  and DNA extractions were then carried out as in [M. marinum-specific primers using a standard curve with previously determined bacterial loads as described in [To assess vaccine efficacy on a primary infection, AB fish immunized with experimental or control (GFP) antigens (~15 fish/group) were infected with a low dose (~40 cfu) of ut as in . The bacM. marinum and followed for twelve weeks. Fish included in the survival experiments were monitored daily for their well-being and humane end point criteria ratified by the national ethical board were followed. Fish showing signs of discomfort or disease were euthanized using 0.04% 3-aminobenzoic acid ethyl ester. Antigen immunizations that showed a protective effect (or tendency)  were repeated one or two more times with similarly sized groups.For survival experiments, the control and experimental fish (19–34 fish/group) were infected with a high dose (~10.000 cfu) of α (1.96) is a constant set based on the accepted error α (0.05), Z1-β (0.8416) is a constant set according to the power of the study (0.8), σ is the estimated standard deviation (0.5). Δ is the difference in the effects of the two treatments compared (estimated effect size), and was set to 0.5 (50%) relating to a reduction in the bacterial burden or improvement in the survival percentage. This is approximately the same as the effect that is achieved by the BCG vaccination [The required sample size (n) for each experiment was calculated using the following formula: cination ,39. Basecination .Statistical analyses were done using the GraphPad Prism 5.02 software . The statistical tests used were the log rank Mantel–Cox test for the survival experiments, and the Mann–Whitney test for bacterial counts and ELISA results. Values of p≤0.05 were considered significant.For the vaccine antigen screen, we selected genes that belong to diverse functional categories and are expressed at different stages of the mycobacterial life cycle. In addition, we chose antigens with different (observed or predicted) cellular locations, although we focused on secreted and membrane-associated proteins, as these presumably are more likely to elicit responses by the host immune system . We chosRpf deficiencies display defects in replication, reactivation and in persistence to stress, presumably due to alterations in the structure of their cell wall [rpf genes in the M. tuberculosis genome (rpfA-E) [M. marinum  homologous to M. tuberculosis rpfC, which is hereafter referred to as rpfC). Despite the name, the expression profiles of the M. tuberculosis rpf genes differ according to the infection phase, suggesting that they have distinct functions [M. marinum antigens in our screen.Resuscitation promoting factors (Rpf) are proteins with peptidoglycan-hydrolysing activity and are thought to be important for mycobacterial virulence and especially for resuscitation from dormancy. Mutant bacterial strains with ell wall ,51–53. T(rpfA-E)  and fourunctions . As Rpfsunctions  and in hunctions ; for a runctions , we inclpe/ppe genes constitute ∼10% of the genome of pathogenic mycobacteria, and their expression is differentially regulated by stress and other environmental conditions, including inside granulomas. PE/PPE proteins are commonly localized to the bacterial cell surface or are secreted, enabling them to elicit and modulate host immune responses. Many PE/PPE proteins have been shown to be highly antigenic [Another relatively well-studied group of potential mycobacterial antigens are the PE/PPE proteins, which are named after the proline-glutamic acid (PE) and proline-proline-glutamic acid (PPE) motifs near their N-termini. The ntigenic  and seventigenic . For ourM. tuberculosis induces strong IFN-γ responses in cattle [M. tuberculosis, Rv3620c, is a secreted, antigenic protein [M. tuberculosis has been shown to induce the activation of memory T cells in humans [M. tuberculosis [We also included proteins with a signature expression profile in certain phases of the infection. For this, we chose the outer membrane protein A (ompA), whose homolog in n cattle , and then cattle . In addin humans ; and MMArculosis ,41.M. tuberculosis homolog Rv2289 shows high abundance in the virulent H37Rv strain, but is nearly absent from the avirulent H37Ra strain [Knowing that mycobacteria undergo extensive metabolic changes during the different stages of their lifecycle , we chosa strain .M. tuberculosis homologs are listed in The selected antigens together with their M. marinum genes in the ATCC 927 strain was verified by qRT-PCR (RpfA and esxM) to 10−5 (pe31) compared to the reference gene MMITS expression, each of the selected genes was verified to be expressed. Based on this, all 15 antigens were selected for further studies and cloned into an expression vector as GFP–tagged fusion proteins.The expression of the selected  qRT-PCR . For thiOne of the key issues in DNA vaccination is achieving adequate antigen expression in the target tissue. To assess this, the in vivo expression of the vaccine constructs was analyzed with three different methods, each utilizing the GFP tag fused with the antigen. First, antigen expression was visualized in situ, in the dorsal muscles of the fish, with a fluorescent microscope. Although not quantitative, visual inspection provides a quick and easy way to evaluate successful vaccinations and antigen expression in each individual fish without harming them. As shown in To quantify antigen expression, a GFP enzyme-linked immunosorbent assay (ELISA) was used for proteins extracted from the dorsal muscles of the vaccinated fish. The quantitated expression of the recombinant constructs relative to the GFP control (samples from fish injected with an empty plasmid) is shown in To validate the correct size of the recombinant antigen fusion proteins, they were visualized with Western blotting, using a HRP-conjugated GFP antibody . The GFPM. tuberculosis infection most often leads to a sub-clinical, latent infection, where the infection retains the potential to reactivate [M. marinum infection, which in most fish leads to a latent disease with stable bacterial counts [M. marinum. Five weeks post-infection, the fish were sacrificed and the bacterial burden of each fish was quantified by qPCR  and PE13 (Rv1195), have been studied using a recombinant M. smegmatis strain. Both recombinants led to the enhanced survival of bacteria within macrophages, presumably due to interference with host (innate) immune signaling pathways [pe13 was upregulated by diverse types of stress, and led to the increased production of interlukin-6 (IL-6) and IL-1β in macrophages [Four antigens were found to have protective effects against a low-dose mycobacterial infection. These include the probable CDP-diacylglycerol pyrophosphatase cdh, and two antigens belonging to the PE/PPE family, namely PE5_1 and PE31, and RpfE. Of these, cdh remains rather poorly characterized. Both the PE5_1 and PE31 antigens led to an approximately 50% reduction in the median cfu counts compared to the control group in the low-dose pathways ,78. The rophages , while Pex vivo model, RpfE induced the maturation of dendritic cells via the TLR4 leading to the generation of Th1 and Th17 cell mediated immunity, without stimulating the suppressive regulatory T cells [M. tuberculosis both in terms of cfu burdens and survival times [M. marinum infection in adult zebrafish displays a natural latency that can be reactivated experimentally or spontaneously [M. tuberculosis carriers, especially adolescents and young adults, who are the main source of TB transmission, would effectively limit new infections [M. marinum infection by the BCG vaccine [Of the mycobacterial antigens included in our screen, the Rpf proteins are probably the best studied, both considering their role in mycobacterial pathogenicity and their potential medical use. The latter is supported also by the results of this study, where RpfE was the only antigen that provided protection against both a primary (low dose) and a fulminant (high dose) infection. This is in line with previous results from mouse studies. In a mouse  T cells . RpfE haal times ,80. Consaneously , the zebfections . We have vaccine ,39, and M. marinum infection model in the adult zebrafish is suitable for early-stage pre-clinical TB vaccine screening and that the PE/PPE proteins and Resuscitation promoting factors, especially RpfE, are interesting candidates for further studies as antigens for DNA vaccines against TB.In conclusion, this study indicates that the S1 FigM. marinum infection (~40 cfu). Five weeks post infections, fish were euthanized and their internal organs were collected for DNA extractions. Bacterial burdens were determined from the extracted DNAs by qPCR with M. marinum specific primers. Figures show the pooled results of different experiments, which are indicated with different colors. Each dot represents the bacterial count in one fish, and the horizontal lines represent median values. N = 10–29. * p<0.05 (two-tailed Mann-Whitney test). Abbreviations: PE5, PE5_1; PE_19, PE19_1; 4207, MMAR_4207; 3501, MMAR_3501.AB zebrafish were vaccinated intramuscularly with experimental antigens and a control (GFP), followed by an intraperitoneal (TIF)Click here for additional data file.S1 Table(DOCX)Click here for additional data file.S2 Table(DOCX)Click here for additional data file.


The repertoire of free-living protozoa in contact lens solutions is poorly known despite the fact that such protozoa may act as direct pathogens and may harbor intra-cellular pathogens.Between 2009 and 2014, the contact lens solutions collected from patients presenting at our Ophthalmology Department for clinically suspected keratitis, were cultured on non-nutrient agar examined by microscope for the presence of free-living protozoa. All protozoa were identified by 18S rRNA gene sequencing.. Acanthamoeba amoeba in 16 solutions (80 %) collected from 12 patients and Colpoda steini, Cercozoa sp., Protostelium sp. and a eukaryotic more closely related to Vermamoeba sp., were each isolated in one solution. Cercozoa sp., Colpoda sp., Protostelium sp. and Vermamoeba sp. are reported for the first time as contaminating contact lens solutions.A total of 20 of 233 (8.6 %) contact lens solution specimens collected from 16 patients were culturedThe repertoire of protozoa in contact lens solutions is larger than previously known. Acanthamoeba have been documented in CL solutions [Contact lens (CL) wearers are at risk of developing infectious keratitis . In partolutions , 4, 5.Here, we prospectively search for free-living unicellular protozoa in CL solutions collected from patients with suspected keratitis, in an effort to broaden the repertoire of free-living protozoa as potential cornea pathogens.Enterobacter aerogenes. The non-nutrient agar plate was incubated at 28 °C in a humidified atmosphere (contact with moistened gauze) and examined by microscope at × 4 and × 10 magnifications. Free-living protozoa were subcultured on a new non-nutrient agar plate with living E. aerogenes in order to obtain sufficient clonal populations. When the growth was sufficient, areas where protozoa were easily detected by microscope were cut and centrifuged at 2000 g for 10 min. The pellet was re-suspended in 1 mL of Amoeba Page’s saline PAS  for further DNA extraction.CL solution specimens were collected between 2009 and 2014 by CL wearers presenting to the Ophthalmology Department of the Timone Hospital in Marseille, France, for the clinical diagnosis of keratitis and corneal ulcers. Clinical criteria for diagnosis included evidence of a corneal infiltrate or corneal ulcer with underlying inflammation, which could lead to the necrosis of corneal tissue. CL solution provided by the patient was poured into a sterile can kept at room temperature for 4–24 h before it was analysed in the laboratory. The following standard protocol was used to search for protozoa. The CL solution was spread onto a non-nutrient agar plate, plated with a lawn of living 2 atmosphere. For the culture of yeasts and fungi, CL solution specimens were seeded onto Sabouraud agar containing chloramphenicol and gentamicin (bioMérieux), incubated at 32 °C for 10 days. All the bacterial isolates were identified using matrix-assisted laser desorption ionization-time of flight mass spectrometry  as previously described [CL solution specimens were seeded onto 5 % sheep-blood agar  and BCYE  and incubated at 32 °C for 10 days in a 5 % COescribed . BrieflyCandida albicans DNA. Sterile distilled water was used as a negative control. PCR consisted of a 15-min initial denaturation Taq polymerase Hot-Star at 95 °C followed by 30-s denaturation at 95 °C, 30-s hybridation at 60 °C and 1-min elongation at 72 °C. After 35 cycles, extension was performed for 5 min at 72 °C. Amplified products were visualized under UV illumination with Syber Safe ® staining after electrophoresis using a 1.5 % agarose gel. PCR products were cloned by the pGEM® -T Easy Vector System Kit according to the manufacturer’s instructions . They were sequenced in both directions using the Big Dye® Terminator V1.1 Cycle Sequencing Kit (Applied Biosystems). Original sequences have been submitted to GenBank.Total DNA was extracted using the QIAmp tissue kit according to the manufacturer’s protocol . A 328-bp fragment of the 18S rRNA gene was PCR-amplified using the primers NS5/F 5′AACTTAAAGGAATTGACGGAAG3′ and NS6/R 3′GCATCACAGACCTGTTGCCTC5′ and an annealing temperature of 60 °C . All ampwww.ncbi.nlm.nih.gov). The highest percentage of sequence similarity was used to identify isolates. Sequence similarity higher than 97 % with a described species was considered to be indicative of identification at the species level. Phylogenetic analysis was established by the neighbor-joining method using MEGA5 software (www.megasoftware.net). Phylogenetic construct was based on the 18S rRNA gene sequences aligned with 52 references.Sequencing products were resolved using an ABI PRISM 3130 automated sequencer (Applied Biosystems). Sequences were compared with the GenBank database using the online BLAST program (Acanthamoeba in 16 (80 %) solution specimens collected from 12 different patients, Colpoda steini in specimen n°14, Cercozoa sp. in specimen n°12, Protostelium sp. in specimen n°15, and an identical 99 % sequence similarity with both Hartmanella and Vermamoeba genus in specimen 13.A total of 20/233 (8.6 %) CL solution specimens collected between 2009 and 2014 from 16 patients, cultured at least one free-living protozoa  specimens, followed by Klebsiella sp. in 4/20 (20 %) specimens, Aeromonas sp. in 3/20 (15 %) specimens, Chryseobacterium sp. and Sphingobacterium sp. in 2/20 (10 %) specimens and Achromobacter sp., Agrobacterium sp., Alcaligenes sp., Citrobacter sp., Delftia sp., Enterobacter sp., Microbacterium sp., Mycobacterium sp., Raoultella sp., Serratia sp., Shewanella sp. and Wautersiella sp. in 1/20 specimens. Fungi were cultured in five protozoa-positive CL specimens. Fungi included Candida guilliermondii, Candida parapsilosis, Candida lipolytica and Candida colliculosa, Fusarium oxyoprum, Sacrocadium kiliense and Penicillium chrysogenum. In three cases, several fungi were co-cultured, including P. chrysogenum, C. parapsilosis and F. oxysporum in case 16–1, C. guilliermondii and F. oxyoprum in case 7 and C. parapsilosis and C. lipolytica in case 10.Twelve of the 20 protozoa-positive (60 %) CL specimens cultured bacteria, while eight protozoa-positive CL specimens did not. Acanthamoeba amoeba. A previous study reported 28 Acanthamoeba isolates from CL solutions, including A. lugdunensis, A. hatchettii and A. castellani [A. rhysodes, A. polyphaga and A. hatchetti were reported in CL specimens of patients with clinical keratitis in Austria [Acanthamoeba are known to cause keratitis [We embarked upon a prospective study of the repertoire of free-living protozoa in the CL solutions. In this study we observed that, unsurprisingly, the vast majority of positive specimens grew an stellani . Furtherstellani . Also, aeratitis –12.Colpoda sp., Protostelium sp., and Vermamoeba sp. in these CL solutions. Likewise, we found no cases of keratitis which were due to any of these three species: non-Acanthamoeba keratitis were found to be due to Valkampfia and Hartmanella amoeba [However, we failed to find a amoeba , 14.Acanthamoeba in particular, have been shown to host so-called amoeba-resisting bacteria [Acanthamoeba, are amoeba-resisting bacteria, including P. aeruginosa [Mycobacterium sp. [Aeromonas sp. [Cercozoa sp., Vermamoeba sp. and Protostelium sp., but not with C. steini, suggesting further studies of the relationships between these protozoa and bacteria may be required.Amoeba, and bacteria , 16, makbacteria . Severalruginosa  Mycobactrium sp. –21 and Aonas sp. . We alsoIn conclusion, the spectrum of protozoa contaminating CL solutions is broader than previously thought. These protozoa may also host ocular pathogens including bacteria and fungi. Some of these emerging protozoa escape the current routine detection of amoeba in clinical specimens collected from corneal lesions, underscoring the need to develop additional laboratory tools for the diagnosis of keratitis.