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Educational policy analysis archives.
n Vol. 14, no. 4 (February 06, 2006).
Tempe, Ariz. :
b Arizona State University ;
Tampa, Fla. :
University of South Florida.
c February 06, 2006
Legend of the large MCAS gains of 20002001 / Gregory Camilli [and] Sadako Vargas.
Arizona State University.
University of South Florida.
t Education Policy Analysis Archives (EPAA)
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Readers are free to copy display, and distribute this article, as long as the work is attributed to the author(s) and Education Policy Analysis Archives, it is distributed for noncommercial purposes only, and no alte ration or transformation is made in the work. More details of this Creative Commons license are available at http:/ /creativecommons.org/licen ses/by-nc-nd/2.5/. All other uses must be approved by the author(s) or EPAA EPAA is published jointly by the Colleges of Education at Arizona State University and the Universi ty of South Florida. Articles are indexed by H.W. Wilson & Co. Send commentary to Casey Cobb (c email@example.com) and errata notes to Sherman Dorn (epaa-editor@s hermandorn.com). EDUCATION POLICY ANALYSIS ARCHIVES A peer-reviewed scholarly journal Editor: Sherman Dorn College of Education University of South Florida Volume 14 Number 4 February 6, 2006 ISSN 1068Â–2341 The Legend of the Large MCAS Gains of 2000Â–2001 Gregory Camilli Sadako Vargas Rutgers, The State University of New Jersey Citation: Camilli, G. & Vargas, S. (2006). The legend of the large MCAS gains of 2000Â– 2001. Education Policy Analysis Archives, 14 (4). Retrieved [date] from http://epaa.asu.edu/epaa/v14n4/. Abstract Issues related to student, teacher, and school accounta bility have been at the forefront of current educational policy initiatives. Recently, the state of Massachusetts has beco me a focal point in debate re garding the efficacy of highstakes accountability models based on an ostensibly large gain at 10th grade. This paper uses an IRT method for evaluating the validity of 10th grade performance gains from 2000 to 2001 on the Massachu setts Comprehensive Assessment System (MCAS) tests in English Language Arts (E LA) and mathematics. We conclude that a moderate gain was obtained in ELA and a small gain in mathematics. Keywords: MCAS; Performance Standards; Item Response Theory; Validating Score Gains; Accountability. Introduction Apparent achievement gains on high-stakes tests have been received with mixed reviews. Some researchers hold a positive view of the potenti al of high-stakes graduation tests while others remain unconvinced of the positive impact of high -stakes designations. For example, Mehrens and Cizek (2001) argued that Â“increases in scores mo st often represent real improvement with respect to the domain the tests sampleÂ” (p. 481). Koretz, Linn, Dunbar & Shepard (1991), on the other hand,
Education Policy Analysis Archives Vol. 14 No. 4 2 compared test results on an existing third-grade test to a newly introduced high-stakes test. They found that average student scores rose for the ensuing four years on forms of the new test. However, in the sixth year students showed essentially no growth when reassessed with the original test. More recently, a number of reports have prov ided widely different and inconsistent conclusions regarding student achievement in regard to the No Child Left Behind law (Braun, 2004; Education Trust, 2004; Fuller, 2004 ; Webb & Kane, 2004). Resolving inconsistencies in results from high-s takes accountability efforts is a key challenge in educational research because tests have substantial consequen ces for test-takers, parents, and educators. It is important to know how well the accountability models (and associated high-stakes testing) are working in order to support the cl aim that they are legitimate tools for driving educational reform. In this paper, we argue that measurement data used to analyze trends in test scores must also be scrutinized because such data lie at the very core of claims of efficacy or invalidity. Increases in test scores may have vari ous causes, and to facilitate analyses for sorting through competing hypotheses, we must be sure that the methods by which tests are created, scored, and maintained do not lead to false impressions of achievement trends. In 2001, students taking the 10th grade test of the Massachusetts Comprehensive Assessment System (MCAS) obtained a large gain over students of the previous yearÂ’s administration. This gain was consequently taken as strong evidence by some observers of the efficacy of high-stakes accountability based on test results. In this paper, we examine the validity of the 2000Â–2001 gain. As we show below, rather than large gains, there wa s a moderate gain in English Language Arts (ELA) and a small gain in 10th grade mathematics. We conclude th e paper with a discussion of why the consequences of the errors in estimating studen t gains in Massachusetts have resulted in a mixed blessing. Policy Context The 2000Â–2001 MCAS gain in Massachusetts is particularly important because it has been taken as prima facie evidence of the efficacy of high stakes consequences. In the spring of 2001, 81% of high school sophomores in Massachusett s passed the ELA subject area of the MCAS examination. Similarly, 75% of high school sophom ores passed the Mathematics subject area of the MCAS examination. The ELA and Mathematics pass rates for 10th graders in 2001 suggest sizable performance improvements from the previous yearÂ’ s assessment in which pass rates were 66% and 55%, respectively. Here, the pass rate is the percen tage of students in the performance categories Advanced Proficient and Needs Improvement that is, the percentage of students who did not fall in the fourth and final category, which is labeled and Failing or Warning A more complete set of these statistics is given in Table 1. Table 1 Percentage of Tenth Grade Students Passing MCAS ELA and Mathemat ics Assessments, 1998Â– 2004 Subject 1998 1999 2000 2001 2002 2003 2004 Mathematics 58 (24) 57 (24) 55 (33) 75 (45)75 (44)79 (51) 85 (57) English 72 (38) 68 (34) 66 (36) 81 (50)86 (59)89 (61) 90 (62) Passing means above the Failing category. Combined percentage for Proficient and Advanced is in parentheses for each year and subject.
The Legend of the Large MCAS Gains of 2000Â–2001 3 The corresponding percentages of students achieving at the Advanced and Proficient levels are shown in parentheses in Table 1. It can be noted that there is relatively no trend for the years 1998Â–2000 and then a sharp jump in 2001 when passing these two tests became a requirement for graduation Increases were very consistent across Racial/Ethnic groups, and were fairly consistent across high schools. According to Michael Russell and Laur a OÂ’Dwyer from Boston College (personal communication), the large increases in both 10th grade ELA and Mathematics scores have been attributed by the Massachusett s Department of Educati on to an increase in student motivation in preparation and performance as well as to improvem ents in the quality of instruction. The 2000Â–2001 increase has been received wi th warm enthusiasm by policy makers. Finn (2002) viewed the 2000Â–2001 gain as a signal that testing, as a component of accountability, functions to increase student learning: On Monday, Department of Education offici als released the resul ts of the spring 2001 MCAS exams which show ed that 82% of 10th graders passed the English test and 75% passed the math testÂ—inc reases of 16% and 20%, respectively, from the previous year. The resultsÂ—which would be good news at any timeÂ— are all the more pleasing because high school studen ts must now pass these sections of the MCAS to graduateÂ…. Now that it has teeth [emphasis added], the MCAS is even better poised to promote reform and boost student achievement. Likewise, Cizek (2003) touted the MCAS testing program as pr oviding Â“strong evidence of positive consequencesÂ” of high stakes testing (p. 42). Regarding positive consequences of testing in Massachusetts, he argued that th e glass is more than half full: The combination of real in creases in learning, gap decrease, student motivation, and drop-out prevention makes this resul t the equivalent of the big E on a Snellen chart. It seems possible to discount this as evidence that tangible positive consequences are actually ac cruing in the context of hi gh-stakes testing only if one is not even looking at the glass. (p. 43) Others have argued that the test scores must be further examin ed, but have accepted the basic validity of MCAS gains. Gaudet reported that even with the large overall gain, in 2002 only about 70% of students in th e poorest communities passed the MCAS after retesting: Looking at the overall numbers, howev er, gives us no information about the specific challenges that face us. While it is encouraging that the MCAS pass rate increased dramatically between 2000 and 2001, there are still many students who have not yet mastered the ba sic skills need ed to live and work in contemporary Massachusetts. (Gau det, 2002, p. 2) It seems clear that the MCAS has become the ce ntral and unquestioned means of describing the success of educational practices in Massachusetts, and continues to receive intense media focus. MCAS results for 2000 and 2001 We used a method based on item response theory (IRT) for evaluating the 2000Â–2001 change in test performance on the 10th grade MCAS. In particular we examined the large gain based percentage of students who passed the test, i.e., ha d scores above a particular criterion score. The methods we used directly estimated the distribution of scores assuming normality. Technical details concerning the distributional methods are given in the Appendix. Here we only note that the basic approach has several advantages: it is unaffected by rounding, it accounts for measurement error, it
Education Policy Analysis Archives Vol. 14 No. 4 4 avoids complexities due to inestimable proficienc ies for individual students, and it is broadly applicable to programs using IRT technologies. Statistical descriptions of MCAS 2000 and 20011 Classical and IRT statistics are given in Tables 2 and 3 for operational 10th grade MCAS items in 2000 and 2001. Also included are the cut scores for the Warning ( Failing ) achievement band. The p -values indicate success rates on both multip le choice (MC) and open response (OR) items for ELA were higher in 2001. The more interest ing statistics concern the relative difficulties of the test items in 2000 and 2001, which ca n be determined by examining the IRT b -values which are estimated in a way that is comparable across years by means of test equati ng (fixed common item parameter or FCIP equating is used). The average IRT b -values indicate that the difficulty of the MC items stayed about the same, and the difficulty of the OR items decreased dramatically. The b -values roughly follow the scale of z scores (with a mean of 0.0 and standard deviation of 1.0), and lower b s indicate easier items. We refer to the scale of the b s as the Â“logitÂ” scaleÂ—where the term logit is derived from the IRT Â“logisticÂ” item model. In IRT, estimates of student ability (also termed proficiency), labeled also have this logit scale. Based on average b -values, the test became easier in 2001 But in Table 2, it can be seen that the cut score (39.0) in 2001 was two points lower than the cut score for 2000 (41.0). This is an unusual finding, even with one less MC item in 2001. Under normal circumstances, if a test gets easier, the cut point would be expected to go up to remain consistent with the previous yearÂ’s test. The fact that the cut point dropped is curious. Table 2 Key Test Characteristics, MCAS Engli sh Languages Arts Test, 2000 and 2001 2000 2001 Item Type Items p -value Average b p -value Average b Multiple choice* (MC) 36 .70 -0.74 .74 -0.74 Open response (OR) 6 .54 -0.40 .61 -1.85 Raw score cut point 41 39 One MC item was dropped in 2001. Source: Massachusetts Department of Education, 2002a, 2002b. In Table 3, it can be seen that the p -values indicate that for Mathematics, success rates on multiple choice (MC), short answer (SA), and open response (OR) items were also slightly higher in 2001 than 2000. The average IRT b -values indicate that the average difficulty for the 38 MC and OR items dropped, though the difficulty of 4 SA item s increased slightly. Thus, the Mathematics test also became substantially easier, yet it can be seen in Table 3 that the cut score (20.0) in 2001 was one point lower than the cut score for 2000 (21.0). This finding is even more curious than that for ELA. 1 More information for exam ining the 2001 MCAS phenomenon ca n be found in the MCAS technical reports available from the Massachusetts Department of Education website.
The Legend of the Large MCAS Gains of 2000Â–2001 5 Table 3 Key Test Characteristics, MCAS Engli sh Languages Arts Test, 2000 and 2001 2000 2001 Item Type Items p -value Average b p -value Average b Multiple choice* (MC) 32 .49 0.51 .55 0.43 Short answer (SA) 4 .43 0.27 .44 0.36 Open response (OR) 6 .35 0.63 .52 0.29 Raw score cut point 21 20 One MC item was dropped in 2001. Source: Massachusetts Department of Education, 2002a, 2002b. For both Math and ELA, OR items contribute about one-half of the total possible points on the assessment. And because the OR items will be come an important part of the investigation reported in this paper, we will examine changes in scoring procedures for the OR items in the next section before returning to an analysis of th e apparently problematic Â“FailingÂ” cut score. Possible MCAS 2000Â–2001 Scoring Changes The 6 OR items from the MCAS Mathematics and their scoring rubrics from 2000 and 2001 are given for public examination on the Massachus etts DOE website. Each OR item was scored on a scale 0Â–4 and the abbreviated descriptions of the rubric description for scores of 4 and 2 are given in Tables 4Â–5. In Table 4, the left-hand column co ntains 2 numbers that the item identifiers for the years 2000 and 2001, respectively. Looking down the second column, it is evident that to obtain a score of Â“4Â” in 2000, the adjective s Â“correctÂ” and Â“accurateÂ” appear for each of the 6 items, while the word Â“correctÂ” appears just twice in 2001. Moreover, the 2000 rubrics appear semantically denser than those in 2001, and a similar pattern ex ists for the score of Â“2.Â” These observations are quite consistent with the 2000Â–2001 rubric s for ELA items scored on the 0Â–4 scale. Table 4 MCAS ELA Â“4Â” Rubric Descript ors, 2000 and 2001 Tests Released Item (2000, 2001) 2000 descriptors 2001 descriptors 13, 13 accurately using; communicate correct strategy facility with 16, 16 accurately creating and interpreting correctly analyzing; using & explaining correct 21, 21 correct procedures; consistent accuracy 22, 22 determining correct; accurately solving 41, 40 accurately applying correct communicates 42, 41 correctly solving; developing describing & analyzing
Education Policy Analysis Archives Vol. 14 No. 4 6 Table 5 Analysis of rubric descriptor differences required for a score of 2 on the scale 0Â–4. Released Item (2000, 2001) 2000 descriptors 2001 descriptors 13, 13 basic understanding; accurately applying; communicating correct partial understanding 16, 16 general understanding; applying Â… with some accuracy; creating an accurate basic understanding; analyze 21, 21 some understanding; implementing correct procedures at least once basic understanding 22, 22 basic understanding; correct strategy in solving; making one or more correct partial understanding 41, 40 general understandin g; making some accurate Â…; applying some correct basic understanding; communicates 42, 41 basic understanding; using appropriate strate gies to solve problems; developing basic understanding; describing Although one might be tempted to attribute the 2000Â–2001 increases in student performance to the OR items, it is also the case that the equated IRT b parameters, which describe item difficulty, should take this fact into account. 2 All things being equal, a simple change in item difficulty will not affect the percentage of students that reach a particular achievement level on the MCAS. As we shall show below, however, it does appear plausible that some of the change in difficulty of OR items may not be accounted fo r by the IRT scaling and equating process. Methods Data and Sample Description A copy of the MCAS 2000 and 2001 10th grade data were obtained without student, school, or district identifiers.3 Students who were classified as LEP ( n = 846) or who had raw scores of zero were not included. This resulted in population sizes of n = 57,542 for ELA 2000, n = 61,968 for 2 Â“The item calibration for the 2001 and 2000 groups was performed separate ly using the combined IRT models (three parameter logistic [3PL] for multiple choice items, two parameter logistic [2PL] for short answer items, and the graded response model [GRM] for open-response items). Calibration of parameter estimates in 2001 placed items on the same scale as in the 2000 calibration by fixing the parameters for the anchoring items to 2000 calibration values. It is noteworthy that at least 25 percent of the 2001Â–2000 equating items were also used for 2000Â–1999 equating, so that their parameters we re actually fixed to 1999 calibration valuesÂ” (M DOE, 2002a, p. 48). 3 Data were provided by researchers at the Ce nter for the Study of Testing, Evaluation and Educational Testing at Boston College. The data obtain ed were strictly anonym ous. No information was present in the data file regarding student, teacher, school, or district identity.
The Legend of the Large MCAS Gains of 2000Â–2001 7 ELA 2001, n = 59,946 for Mathematics 2000, and n = 62,900 for Mathematics 2001. These values are relatively close to the sample sizes n = 57,681, 62,620, 59,978, and 62,921 given in the 2000 and 2001 technical reports (Massachusetts Department of Education [MDOE], 2002a & 2002b) for the classical reliability statistics. Data consisted of responses for each student to the 42 ELA and Mathematics operational items in 2000 and the 41 ELA and Mathematics operational items in 2001. Cut Score Drift The primary source of bias investigated in this paper concerns changes in IRT scaling procedures. A series of complex changes occurred from 2000 to 2001 that impacted both scale and cut scores. In particular, the reporting scale scores which had been linked to the original raw score metric from 1998 to 2000, were modified in 2001.4 For quality control purposes, the contractor conducted an investigation in which the implemented cut score was examined for potential drift. In particular, the cut score was examined in terms of its IRT equivalent by year. For 10th grade, the relevant graph in the 2001 MCAS technical report (MDO E, 2002a) is given in Figure 1. It can be seen that the cut score in the logit metric appeared to rise in 1999 and 2000 and then drop again in 2001. Figure 1 Tenth grade cut scores mapped against by year, excerpted from the 2001 technical report (MDOE, 2001a, p. 81). 4 The new scale was based on two considerations. First, it was to be based dir ectly on the underlying IRT logit ( ) metric. Second, it was adjusted in a way that, on the surface, resulted in more sensitivity to changes at the lower end of the proficiency continuum. Th e actual procedure is quite complex and is likely to be understood only by psychometricians. However, the results of the rescaling can be analyzed independently of the procedure itself.
Education Policy Analysis Archives Vol. 14 No. 4 8 The legend from the 2001 technical report (MDOE, 2002a) for Figure 1 provides some helpful elaboration: Theta scale (ability or mea sured construct level) established by calibration of the reference forms in 1998. Dashed lines represent cu t points Â… obtained by standard setting in 1998Â…. (p. 81) In 1998 passing scores were set for 10th grade MCAS mathematics and ELA. As new forms of the test are given, these two cut points should remain the same in the logit metric (labeled Â“Ability LevelÂ” in Figure 1) b ecause each new form is equated to the 1998 base year with the FCIP method. The logic is that a cut point is lik e a hurdle, and the bar m ust remain at the same height to accurately gauge passing performance. Th us, the cut point in the logit metric should be a flat line in Figure 1, but it is not. Rather, it can be seen that the cut poin t for the Â“FailingÂ” level drifted upward from its or iginal value of -0.19 in 1998 to -0.06 in 2000. It then sharply decreased from -0.06 to -0.39 in 2001.5 In other words, the cut point dropped about one third of a standard deviation assuming th e logit scale was established as z -score scale in 1998 (a very common IRT practice). The effect of this down ward change (i.e., lowering the bar) for Mathematics, as we shall see in the next section, is much greater than the 2001 technical report estimate of about 2% (MDOE, 2002a, p. 26). For ELA, the cut point for the Â“FailingÂ” level drifted down slightly from its or iginal value of -0.41 in 1998 to about -0.42 in 2000. It then decreased from -0.42 to -0.59 in 2001. Note that pushing the cut point (i.e., the bar) down is equivalent to pushing the entire proficiency distribution up Because the original cut points should not in principle change across administrations, they provide highly useful numerical values for evaluati ng pass rates for subsequent years in the IRT logit metric. We conduct a formal IRT analysis in the next section. However, the effect of the change in scaling is striking when the raw score distribution for mathematics given in Figure 2 is compared to its corresponding MCAS scale score6 distribution given in Figure 3. It can be seen that while raw scores roughly follow a bell-shaped distribution, the scale scores spike at 220 and 260, the cuts for the passing and Proficient levels, respectively (i.e., the lowest scores in the passing and Proficient categories). Visually, it appears that the 2001 rescaling has pushed a number of scores up to the next level. We will show below that this impression is correct. 5 The -0.39 value appears in Table 126.96.36.199 on p. 25 of the 2001 technical report (MDOE, 2002a). The value -0.06 was approximated from Figure 1. 6 Currently, student scores are first defined in th e IRT logit metric, which requires decimal values, and then transformed for practical pu rposes to a reporting scale of whol e numbers that runs from 200 to 280.
The Legend of the Large MCAS Gains of 2000Â–2001 9 60 50 40 30 20 10 0 MCAS Mathematics Raw Score 1,500 1,000 500 0 Frequency Figure 2 Frequency distribution of 10th grade 2001 MCAS mathematics raw scores, with scores of zero omitted. 280 260 240 220 200 180 MCAS Mathematics Scale Score 7,000 6,000 5,000 4,000 3,000 2,000 1,000 0 Frequency Figure 3 Frequency distribution of 10th grade 2001 MCAS mathematics sc ale scores (with corresponding raw scores of zero omitted).
Education Policy Analysis Archives Vol. 14 No. 4 10 Preliminary IRT Analyses7 As a first step, we attempted with IRT analyses of the 2000 and 2001 examinee data for both Mathematics and ELA to reproduce the operational b parameter estimates (or IRT item difficulties) as reported in the MCAS technical manuals. For the 2000 ELA and Mathematics examinations, we were able to reproduce the item difficulties for a ll items. However, when the same steps were taken with the 2001 test data, we observed large discrepancies between our OR b -values and those reported in the MCAS technical manuals. Recall that as a b -value moves in the negative direction, a test item becomes easier. In 2000, our OR b -values were lower than the reported MCAS va lues by approximately .10 logits for the ELA examination and approximately 0.03 logits for the Mathematics examination. In contrast, our b -values for OR items in 2001 were lower by approx imately 0.20 logits for the ELA examination and 0.40 logits for the Mathematics examination. (Not e that these logit differences can be interpreted accurately as effect sizes). Thus, in 2001, it appears that the OR items were easier than expected relative to the other test items.8 Main Analysis Given this discrepancy, we used the two alterna tive IRT methods (given in the Appendix) to estimate pass rates. The first used OR b -values from the MCAS technical report, and the second adjusted the b -values for the decrease in difficulty from 2000 to 2001. In both methods, the original 1998 cut scores in the logit metric, as shown in Figure 1, were employed. Additional description is given in the Appendix. Results The results of these analyses are given in Table 6, which presents the reported total percentages of students who passed each examinat ion in each year as well as the percentage of students falling in the Proficient category or higher. An Â“adjusted reportedÂ” category is indicates the percentage based solely on students who were present for the examination in each category. When reporting results, those students who were not present for an examination (but were eligible to participate) were considered failing. There is approximately a 2% difference be tween the reported pass rates and the adjusted pass rates in 2000, while less than 1% in 2001. For simplicity, we shall consider the adjusted MCAS pass rates when drawing our conclusions since our samples did not include non-present student (i.e., those students considered failing, not present). 7 We hope that psychometricians will understand that we have ac cess to the operational MCAS b -value estimates of 2000 and 2001. A previous reviewer of this work failed to grasp this elementary point in contending that nothing could be establis hed without access to the linking itemsÂ’ b -values. 8 There are several reasons why this might be the case including pre-equating or samples used for item calibration. We can not determine the exact reason from publicly available information.
The Legend of the Large MCAS Gains of 2000Â–2001 11 Table 6 Reported and Estimated 2001 Percentag es of Passing and Proficient Students English Language Arts Mathematics Parameter 2000 2001 Gain 2000 2001 Gain Reported passing rates Pass 66.0% 82.0% 16.0% 55.0% 75.0% 20.0% Proficient 36.0% 50.0% 14.0% 33.0% 45.0% 12.0% Reported rates, adjusted for non-participation Pass 68.0% 82.8% 14.8% 56.7% 75.8% 19.1% Proficient 37.1% 50.5% 13.4% 34.0% 45.5% 11.4% Official b -values for all items (Method 1) Pass 67.3% 77.8% 10.5% 57.4% 69.3% 11.9% Proficient 37.5% 48.2% 10.6% 34.8% 41.4% 6.6% Official b -values with estimated OR (Method 2) Pass 65.9% 75.7% 9.8% 56.8% 60.9% 4.1% Proficient 35.9% 44.5% 8.7% 34.4% 36.3% 1.9% For the 2000 data, we were nearly able to repr oduce (about a 1% discrepancy) the pass rates9 for both the ELA and Mathematics examinations using the MCAS reported item parameters. The results for 2001 differed by slightly more with this method at approxim ately 5% and 6% for the ELA and Mathematics examinations, respectively However, when we used the 2001 implemented cut scores as shown in Figure 1, we were able to estimate pass rates in 2001 that were within 1Â–2% of adjusted pass rates. The latter finding suggests that potential sample differences did not unduly affect the results. In Table 6, it can be seen that the unad justed pass rate increased from 68% to 82.8% (+14.8%) for 10th grade ELA. Taking into account the drift in the cut score and changes in OR item behavior, we estimated that the adjusted pass ra te increased from 65.9% to 75.7% (+9.8%). Thus, the 2000Â–2001 gain in ELA is likely to be over-est imated by about 5% (14.8% 9.9%). Similarly for Mathematics, we estimated that the adjusted pa ss rate increased from 56.8% in 2000 to 60.9% in 2001. Thus, the 2000Â–2001 gain in Mathematics is likely too high by about 15%. The change in scaling affected all of the MCAS tests to some degree, which can be seen upon inspecting the figures in Appendix I of the 2001 technical report (MDOE, 2002a, Â“MCAS Performance Levels Mapped to Theta ScaleÂ”). We have only analyzed the results for 10th grade ELA and Mathematics in this report; the 2000Â–2001 gains for all 2001 MCAS tests may contain varying degrees of statistical bias. Discussion Our goal is to show how the psychometric aspects of tests (i.e., scaling and equating procedures) can adversely aff ect reported student pass rates.10 States have made large investments in 9 Â“Pass rateÂ” refers to the area of the theta distribution above the reported MCAS theta cut-point for a particular examination. 10 Scaling changes made in 2001 are reported in the Massachusetts DOE on pages 20Â–26 in the 2001 technical manual.
Education Policy Analysis Archives Vol. 14 No. 4 12 assessment programs, and it is critically important that scaling issues do not distort the interpretation of achievement gains. It is important to note at the same time that the present study supports the quality of the MCAS and its technical documentation. Though a flaw was found with the manner in which cut scores were determined, this should not be taken as evidence against the validity or reliability of the test per se. As Cizek (2001) and Popham (2003) have noted, the educational measurement community should be involved in debates regarding the efficacy of testing. However, while Cizek (2001) complained that measurement experts have been silent on the benefits of high-stakes testing, Popham (2003) argued that most students are receiving educations of decisively lower quality as a result of high-stakes testing (in stark contrast to his positions of 20 years ago11). Popham further elaborated that in contrast to sins of commission which are easily spotted, Sins of omission can also have serious consequences. And those are the kinds of sins that the educational measurement crowd has been committ ing during recent decades. We have been silent while certain sorts of assessment tools, the very assessment tools that we know the most about, have been misused in ways that harm children. (p. 46) What sense can be made of these competing claims of Â“s ilenceÂ”? We would argue first and foremost that the co nceptual territory here is not black and white. There are many purposes for testing, and these purposes are ranked differe ntly depending on oneÂ’s values, philosophy, and political ideology. The Massachus etts experience has provided a sort of ink blot into which various beliefs about causality have been projec ted. Measurement speciali sts are not immune to such influences, and there is no reason to believe that they will forge a greater consensus on the issue of high-stakes testing than exists in the prevailin g social context. Howe ver, one thing that all stakeholders can agree upon is the need for accurate estimation of program effects. Yes, there is much explaining to do once the effects ar e measured, but the diff iculty of accurate measurement, especially with regard to annu al progress, should not be underestimated. In Massachusetts it was possible to refute the grander claims regarding the effects of highstakes testing in 2001 because the State Department of Education scrupulously detailed the psychometric characteristics of the MCAS assessment. The kind of analyses undertaken in this paper could not be carried out in most states using publicly available information. This is one reason that we do not currently have a very good notion of how broadly (across states) assessment errors have affected educational policies. Though some errors ha ve been reported, it is likely the case that many others have passed silently (Rhoa des & Madaus, 2003). Many iden tified problems have concerned inconsistent or incorrectly scored items. Such ca ses are relatively easier to detect than those involving scaling, scoring and equating. Yet the la tter are more likely to confuse state educational policies as well as to muddy the debate on the merits of high-stakes testing. Several conclusions and recommendations can be drawn regarding the change in MCAS scores form 2000 to 2001 considered in this paper. First, we did estimate gain s from 2000 to 2001 in both English Language Arts and Mathematics, but the gains were much smaller than those in official reports. The ELA 10th grade gain was moderate resulting in a reported pass rate of 81% in 2001; but in 1998, the 8th grade NAEP rate for Basic and Above (the lowest category being Below Basic ) in reading was 79%. The MCAS gain for Mathematics wa s relatively small resulting in a reported pass rate (partially proficient) of 75 % in 2001; but in 1996, the 8th grade NAEP rate for Basic and Above was 68%. The de facto achievement levels implemented in 20 01 thus appear more consistent with Basic achievement levels set in the National Assess ment of Educational Progress (NAEP) in 8th grade, while the original achievement levels set in 1998 were more severe. Such discrepancies should be understood as general and common problem of setting standards. Standard setting procedures 11 See Camilli, Cizek and Lugg (2001) for a short history of PophamÂ’s views.
The Legend of the Large MCAS Gains of 2000Â–2001 13 are designed to be internally consistent but require essentially establishing arbitrary cut points on a continuum of test scores (Camilli, Cizek & Lugg, 20 01). Different procedures and contexts produce different results, and this topic has remained cont roversial. In 2005, the Ed ucation Department, in a shift from previous practices, presented state resul ts with Â“charts showing state-by-state trends focused on results for just the basic level, which de notes what NAGB regards as Â‘partial masteryÂ’ of the skills students should acquire at particular gr ade levelsÂ” (Viadero & Olson, 2005, p. 14). Critics such as Diane Ravitch have decried this phenomenon as a lowering of standards, while state policy makers tend to view the Basic level as a plausible criterion fo r student proficiency. With NAEP, there has also been controversy regarding how achievement levels should be established and interpreted (Pellegrino, Jones & Mitchell, 1999 ; Hambleton et al., 2000). Nonetheless, the consistency of 2001 10th grade MCAS results with extrapolations from earlier NAEP 8th grade performance might be taken as a positive unintended outcome. Second, the evidence from Massachusetts supporting the efficacy of high-stakes accountability is mixed. A more compelling explanat ion is that mathematics scores had been rising all alongÂ—and the upward trend existed prior to the implementation of new graduation requirements. However, there is evidence from NAEP that reading scores have been stagnant nationally in all grade levels as well as in Massachusetts at 4th and 8th grade. Rising ELA scores in 10th grade thus signify some proficiency not reflected in earlier grades by NAEP. Another hypothesis is that score cut points have continued to drift downward (in the metric) because the method of producing scale scores may be overly sensitiv e to slight changes in testing procedures. Finally, unprecedented gains, such as those whic h occurred in 2001 MCAS proficiency at the 10th grade, should be recognized by scholars as pr ime candidates for further study. Indeed, if an increase in student proficiency seems almost too good to be true, then some degree of skepticism is both appropriate and healthy. Grissmer, Flanagan Kawata, and Williamson (2000) showed that the annual gains of any state on NAEP average about 0.03 (but can be as high as 0.06 ) per year. When a gain nearly an order of magnitude larger than this is observed, as it was in 10th grade MCAS Mathematics, it should receive additional scrutiny This is not simply a technical issue. Failing to obtain accurate estimates of achievement gains can result in false perceptions that lead both educators and pundits astray.
Education Policy Analysis Archives Vol. 14 No. 4 14 References Camilli, G. (1988). Scale shrink age and the estimation of late nt distribution parameters. Journal of Educational Statistics 13 227Â–241. Camilli, G., Cizek, G.J., & Lugg, C.A. (2001). Psychometric theory and the valid ation of performance standards: History and futu re perspectives in G.J. Cizek (ed.) Setting performance standards (pp. 445Â–476), Mahwah, NJ: Lawrence Erlbaum. Education Trust (2004). Measured progress: St ates are moving in the right direction in narrowing achievement gaps and raising ac hievement for all students, but not fast enough. Washington, D.C.: Author. Finn, C.E., Jr. (2001, October 18). Vindication for the MCAS: Dramatic improvement in student scores in MA. The Education Gadfly 1 (22). Retrieved February 1, 2006, from http://www.edexcellence.net /foundation/gadfly/issue. cfm?edition=&id=86#1296. Fuller, B. (2004, October 13). Ar e test scores really ri sing? School reform an d campaign rhetoric. Education Week, 24 (7), 40, 52. Gaudet, R. D. (2002, June). Student achievement in Massachusetts : The lessons of nine years of education reform Education Benchmarks Report #1Â–02. University of Massachusetts, Amherst, MA: Donahue Institute. Grissmer, D.W., Flanagan, A., Ka wata, J & Williamson, S. (2000 ). Improving student achievement: What state NAEP t est scores tell us (MRÂ–924Â–EDU). Sant a Monica, CA: Rand. Koretz, D., Linn, R. L., Dunbar, S. B., & Shepard, L. A. (1991, April). The effects of high-stakes testing on achievement: Pr eliminary findings about generalization across tests Paper presented at the annual meet ing of the American Educati onal Research Association, Chicago, IL. Linn, R.L. (1998). Assessments and accountability. Educational Researcher, 29 (2), 4Â–14. Massachusetts Department of Education. (2001a). Guide to interpreting th e spring 2001 reports for schools and districts Malden, MA: Author. Retrie ved February 1, 2006, from http://www.doe.mass.edu/ mcas/2001/interpretive_ guides/fullguide.pdf. Massachusetts Department of Education. (2001b, November). Spring 2001 MCAS tests: Report of 2000Â–2001 sc hool results Malden, MA: Author. Retrie ved February 1, 2006, from http://www.doe.mass.edu/ mcas/2001/results/sc hool/g10s_0001res.pdf. Massachusetts Department of Education. (2002a). 2001 MCAS technical report Malden, MA: Author. Retrieved February 1, 2006, from http://www.doe.mass.edu/mcas /2002/news/01techrpt.pdf.
The Legend of the Large MCAS Gains of 2000Â–2001 15 Massachusetts Department of Education. (2002b, May). 2000 MCAS technical report Malden, MA: Author. Retrieved February 1, 2006, from http://www.doe.mass.edu/mcas /2002/news/00techrpt.pdf. Mehrens, W.A., and Cizek, G.J., (2001). Standard setting and th e public good. In G.J. Cizek (Ed.) Setting performance standards (pp. 477Â–485), Mahwah, NJ: Erlbaum. Muraki, E., and Bock R. D. (2003). PARSCALE (Version 4.1): Analysis of graded responses and ratings [Computer program]. Chicago, IL: Scient ific Software International, Inc. Popham, W.J. (2003). Seeking rede mption for our psychometric sins. Educational Measurement : Issues and Practices, 22 (1), 45Â–48. Rhoades, K., & Madaus, G. (2003). Errors in standardized t ests: A systemic problem Boston, MA: National Board on Educational Testin g and Public Policy Boston College. Viadero, D., and Olson, L. (2 005, November 2). Focu s on Â‘BasicÂ’ achievem ent level on NAEP stirs concern. Education Week, 25 (10), 14.
Education Policy Analysis Archives Vol. 14 No. 4 16 Appendix Item Calibration All calibrations were carried out using the IRT software program PARSCALE in which MC, SA, and OR items (Muraki and Bock, 2003) can be jointly scaled. A three parameter logistic model was used for MC items; this model was also used for SA items with the guessing parameter set to c = 0. SamejimaÂ’s graded response model was used for OR items. Estimating Pass Rates As a standard feature or PARSCA LE, the posterior distribution of examinee proficiency is output. Also referred to as the latent population distribution (LPD), it has been shown (see Camilli, 1988, for a more extensive discussion) that this distribution is far more accurate than the distribution of estimated abilitie sÂ—especially with respect to measurement error and unestimable examinee proficiencies. Because th e LPD is scaled in terms of IRT item parameters ( a b and/or c ), it is fixed on the basis of the FCIP (fixed common item parameters) equating method used with the MCAS. Both individual examinee item response patterns and item parameter estimates are required to obtain the LPD. A normal prior distribution was used, and Figure 2 suggests this assumption is highly plausible. To use the LPD to estimate pass rates, and original criterion or cut score must be translated into the IRT metric in the year the cut score was determined. For 10th grade, these scores were set in 1998 for both ELA and Mathematics. Once a cut scor e is obtained in the IRT metric, the percentage of the LPD above (or below) the cut is used to es timate the percent above criterion (PAC). For the lowest cut score on the MCAS the PAC is described as the pass rate whereas the percent below is described as the Â“FailingÂ” rate. The LPD is obtained as a series of theta scores (quadrature points) with associated probabilities or Â“weights,Â” that is, as a discrete density function. Percentiles are obtained by summing weights below the cut score, possibly with some interpolation. We found there was little difference between 50 and 100 quadr ature points; all the results below are based on the latter number. Method 1 (fixed MCAS item parameters) Using the item parameters reported in the MCAS technical manuals, we estimated posterior theta di stributions for both examinations in both years using all items (MC, SA, and OR). This method was considered a referent analysis. Though there were minor discrepancies between our samples an d the samples on which the MCAS reports were based, such discrepancies most likely had a negligible effect. Our analyses of the classical item statistics and the IRT item parameters reported in the MCAS manuals for the OR items suggested that some error or combination of errors led to systematic misestimation of these itemsÂ’ parameter s. The results of our Method 1 analyses further supported this belief due to the disproportionately large chi-square misfit statistics observed for these items. More specifically, review of the model-fit outputs for 2001 ELA and Mathematics examinations revealed that OR items statistics (o r rather Â“misfitÂ” statisti cs) were on average twice as large, relative to degrees of freedom, than those observed for the non-OR items. Method 2 discussed below addresses this issue. Method 2 (fixed MCAS MC, estimated OR parameters) Using the procedure discussed above of fixing item parameters, we fixed the MC values on the ELA examination, both the MC and SA values on the Mathematics examination to the reported MCAS values. However, we allowed the OR items to be estimated. Then the posterior theta distributions were again generated for both examinations in both years. The success rates for Pass and Proficient obtained from these analyses (labeled Method 2 in Table 6) were slightly lo wer than Method 1 ELA pass rates. For Mathematics the differences was larger. We estimated success rate s that 7.8% and 4.7%, respectively, less than the Method 1 success rates. This suggests that OR items were more problematic on the Mathematics examination.
The Legend of the Large MCAS Gains of 2000Â–2001 17 About the Authors Gregory Camilli Rutgers, The State Univ ersity of New Jersey Sadako Vargas Rutgers, The State Univ ersity of New Jersey Email: Camilli@rci.rutgers.edu Gregory Camilli is Professor in the Rutgers Grad uate School of Education. His interests include measurement, program evaluation, and policy issues regarding student assessment. Dr. Camilli teaches co urses in statistics and psychometrics, st ructural equation modeling, and meta-analysis. His current research interests include school factors in mathematics achievement, test fa irness, technical and validity i ssues in high-stakes assessment, and the use of eviden ce in determining in structional policies. As Research Associate at Rutgers Graduate School of Ed ucation, and Adjunct Professor at Touro College and Seton Hall University, Sadako Vargas has taught in the areas of research methods and occupational therapy. Her interests li e in the use of meta-analysis for investigating intervention effects in the area of rehabilitation and education specifically related to pediatrics and occupational therapy intervention.
Education Policy Analysis Archives Vol. 14 No. 4 18 EDUCATION POLICY ANALYSIS ARCHIVES http://epaa.asu.edu Editor: Sherman Dorn, University of South Florida Production Assistant: Chris Murre ll, Arizona State University General questions about ap propriateness of topics or particular articles may be addressed to the Editor, Sherman Dorn, firstname.lastname@example.org. Editorial Board Michael W. Apple University of Wisconsin David C. Berliner Arizona State University Robert Bickel Marshall University Gregory Camilli Rutgers University Casey Cobb University of Connecticut Linda Darling-Hammond Stanford University Gunapala Edirisooriya Youngstown State University Mark E. Fetler California Commission on Teacher Credentialing Gustavo E. Fischman Arizona State Univeristy Richard Garlikov Birmingham, Alabama Gene V Glass Arizona State University Thomas F. Green Syracuse University Aimee Howley Ohio University Craig B. Howley Appalachia Educational Laboratory William Hunter University of Ontario Institute of Technology Daniel Kalls Ume University Benjamin Levin University of Manitoba Thomas Mauhs-Pugh Green Mountain College Les McLean University of Toronto Heinrich Mintrop University of California, Berkeley Michele Moses Arizona State University Anthony G. Rud Jr. Purdue University Michael Scriven Western Michigan University Terrence G. Wiley Arizona State University John Willinsky University of British Columbia
The Legend of the Large MCAS Gains of 2000Â–2001 19 EDUCATION POLICY ANALYSIS ARCHIVES English-language Graduate -Student Editorial Board Noga Admon New York University Jessica Allen University of Colorado Cheryl Aman University of British Columbia Anne Black University of Connecticut Marisa Cannata Michigan State University Chad d'Entremont Teachers College Columbia University Carol Da Silva Harvard University Tara Donahue Michigan State University Camille Farrington University of Illinois Chicago Chris Frey Indiana University Amy Garrett Dikkers University of Minnesota Misty Ginicola Yale University Jake Gross Indiana University Hee Kyung Hong Loyola University Chicago Jennifer Lloyd University of British Columbia Heather Lord Yale University Shereeza Mohammed Florida Atlantic University Ben Superfine University of Michigan John Weathers University of Pennsylvania Kyo Yamashiro University of California Los Angeles
Education Policy Analysis Archives Vol. 14 No. 4 20 Archivos Analticos de Polticas Educativas Associate Editors Gustavo E. Fischman & Pablo Gentili Arizona State University & Universidade do Estado do Rio de Janeiro Founding Associate Editor for Spanish Language (1998Â—2003) Roberto Rodrguez Gmez Editorial Board Hugo Aboites Universidad Autnoma Metropolitana-Xochimilco Adrin Acosta Universidad de Guadalajara Mxico Claudio Almonacid Avila Universidad Metropolitana de Ciencias de la Educacin, Chile Dalila Andrade de Oliveira Universidade Federal de Minas Gerais, Belo Horizonte, Brasil Alejandra Birgin Ministerio de Educacin, Argentina Teresa Bracho Centro de Investigacin y Docencia Econmica-CIDE Alejandro Canales Universidad Nacional Autnoma de Mxico Ursula Casanova Arizona State University, Tempe, Arizona Sigfredo Chiroque Instituto de Pedagoga Popular, Per Erwin Epstein Loyola University, Chicago, Illinois Mariano Fernndez Enguita Universidad de Salamanca. Espaa Gaudncio Frigotto Universidade Estadual do Rio de Janeiro, Brasil Rollin Kent Universidad Autnoma de Puebla. Puebla, Mxico Walter Kohan Universidade Estadual do Rio de Janeiro, Brasil Roberto Leher Universidade Estadual do Rio de Janeiro, Brasil Daniel C. Levy University at Albany, SUNY, Albany, New York Nilma Limo Gomes Universidade Federal de Minas Gerais, Belo Horizonte Pia Lindquist Wong California State University, Sacramento, California Mara Loreto Egaa Programa Interdisciplinario de Investigacin en Educacin Mariano Narodowski Universidad To rcuato Di Tella, Argentina Iolanda de Oliveira Universidade Federal Fluminense, Brasil Grover Pango Foro Latinoamericano de Polticas Educativas, Per Vanilda Paiva Universidade Estadual Do Rio De Janeiro, Brasil Miguel Pereira Catedratico Un iversidad de Granada, Espaa Angel Ignacio Prez Gmez Universidad de Mlaga Mnica Pini Universidad Nacional de San Martin, Argentina Romualdo Portella do Oliveira Universidade de So Paulo Diana Rhoten Social Science Research Council, New York, New York Jos Gimeno Sacristn Universidad de Valencia, Espaa Daniel Schugurensky Ontario Institute for Studies in Education, Canada Susan Street Centro de Investigaciones y Estudios Superiores en Antropologia Social Occidente, Guadalajara, Mxico Nelly P. Stromquist University of Southern California, Los Angeles, California Daniel Suarez Laboratorio de Politicas Publicas-Universidad de Buenos Aires, Argentina Antonio Teodoro Universidade Lusfona Lisboa, Carlos A. Torres UCLA Jurjo Torres Santom Universidad de la Corua, Espaa