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Measurement Invariance across groups
Mplus Discussion
Confirmatory Factor Analysis
posted on Tuesday, June 14, 2005 - 8:32 am
Hi there,
I am attempting to produce a final model which takes into account measurement invariance I have found by sex. We are working with a two factor solution, and when we regress the two factors on sex we find significant paths for both factors. Using the modification indices, we then identified several observed variables (loading on factor 2) which show significant paths to sex. When these paths from observed variables to sex were included in the model, the relationship between factor 2 and sex became non-significant. We then took out the path between factor 2 and sex. However, despite the fact that this path was not significant, it had a dramatic effect of the chi-square value, suggesting that the path needed to be in the model. We then added the path back in, but set it to zero. This fixed the chi-square problem. We are unsure as to why this would happen. Why are we getting such dramatically different chi-square values if the paths we remove are non-significant and why does including a path that is set to 0 have such a dramatic effect on the chi-square value?
Thank you so much for you time,
Linda K. Muthen
posted on Tuesday, June 14, 2005 - 9:13 am
Taking a path out and fixing it to zero is the same thing. If you cannot resolve this, you will need to send the relevant outputs and your license number to
Jeremiah Schumm
posted on Wednesday, October 18, 2006 - 8:18 am
I am attempting to compare nested measurement models to test for measurement invariance. I attempted to freely estimate the models paramters for the "female" sample by using the following command:
support BY
combat WITH support;
MODEL female:
support BY
combat WITH support;
However, I am receiving the following warning:THE RESIDUAL COVARIANCE MATRIX (THETA)IN GROUP MALE IS NOT POSITIVE DEFINITE. THIS COULD INDICATE A NEGATIVE VARIANCE/RESIDUAL VARIANCE FOR AN OBSERVED VARIABLE, A CORRELATION GREATER OR EQUAL TO ONE BETWEEN TWO OBSERVED VARIABLES, OR A LINEAR DEPENDENCY AMONG MORE THAN TWO OBSERVED VARIABLES. CHECK THE RESULTS SECTION FOR MORE INFORMATION.PROBLEM INVOLVING VARIABLE FACESCO.
Am I specifying the variant model incorrectly, or does this warning relate some idiosyncracy about the male sample in my dataset? Thank you.
Linda K. Muthen
posted on Wednesday, October 18, 2006 - 10:15 am
There seems to be a problem with the variable facesco. Does this variable have a negative residual variance in the male group? This is usually the problem.
Xuan Huang
posted on Wednesday, May 16, 2007 - 10:03 am
Dear professors:
Could you give us some suggestions on testing measurement invariance in Mplus? We want to test whether parenting measures are equivalent across mothers and fathers. Because the mother and the father are from the same family unit, the two groups in comparison are not independent.
Can we take care of non-independence across groups in multi-level, multi group CFA in MPLus? Thanks a lot in advance.
Linda K. Muthen
posted on Wednesday, May 16, 2007 - 10:15 am
You can do this by taking a multivariate approach where each observation has data for both mothers and fathers. You would then have factors for mothers and factors for fathers and you would place equalities on the measurement parameters to test for measurement invariance. See Example 6.14 which is a growth model and just imagine it without the growth component.
Scott Weaver
posted on Wednesday, May 16, 2007 - 11:03 am
I have a related question. How should one test for measurement invariance of a scale across two groups (defined at Level 1, e.g., male/female) when L1 units are nested in L2 units (e.g., class)?
I think that a multiple-group approach would not be ideal because male students are not independent from female students in the same class - and, to my knowledge, grouping=gender, type=complex, & cluster=class, would only adjust for dependence within each gender group, but not across gender groups. Is this correct?
I also don't think that a multivariate approach would work like it does for the above mother/father scenario. Each observation at the class level would have multiple males and females as opposed to typically a single mother and father at the family level. Maybe aggregrating needs to be done?
I used Mplus to test a multiple-group SEM model with complex data (clustering and stratification). I used type=complex to address the complex nature of the data - but am wondering how dependence of units across groups is handled by Mplus and whether my tests of structural invariance constraints, which I think assume the groups are independent, are biased?
Thanks, Scott
Linda K. Muthen
posted on Wednesday, May 16, 2007 - 1:23 pm
It is true that if clusters contain both males and females, the males and females are not independent groups. With TYPE=COMPLEX in Mplus, an adjustment has been made to take this lack of independence into account.
Scott Weaver
posted on Wednesday, May 16, 2007 - 2:41 pm
So you are saying that Mplus (with type=complex) not only accounts for cluster-based dependence within groups (e.g., males and females), but also cluster-based dependence between gender groups in a multiple group analysis?
Then with Xuan Huang's situation involving mothers and fathers above, is it appropriate (as an alternative to the multivariate approach) to use a multiple group model (group = parent) with type=cluster - which would typically result in a single observation for each cluster within group? If so, should this lead to equivalent results with the multivariate approach you suggested?
Thanks Linda!
Linda K. Muthen
posted on Thursday, May 17, 2007 - 6:25 am
The multiple group approach should yield approximately equivalent results.
Richard E. Zinbarg
posted on Tuesday, July 10, 2007 - 11:56 am
Hi Linda and/or Bengt,
I am having a problem testing for invariance of a second-order CFA with 3 first-order factors and one second-order factor using robust estimation. Using ordinary ML, we encounter no problems though the CFI is lower than what one would like so we wanted to see what the Robust estimate of CFI looked like. When we run the same exact invariance model (across whites vs minorities) we receive a error message saying that the model is not identified due to a problem involving parameter 77. Parameter 77 is in the value in the Alpha vector for the second-order factor in the minority group. Let me know if you need me to send you our input file (and data file).
Rick Zinbarg
Linda K. Muthen
posted on Tuesday, July 10, 2007 - 12:14 pm
So you are saying when you change the estimator from ML to MLR you get the error message?
Richard E. Zinbarg
posted on Saturday, July 14, 2007 - 1:30 pm
Linda K. Muthen
posted on Sunday, July 15, 2007 - 4:00 pm
Please send your input, data, output, and license number to
Alexandre Morin
posted on Friday, May 16, 2008 - 9:56 am
Greetings Linda,
Todd Little and his colleagues propose the "effects coding" method to identify MACS models in various papers. This method involves constraining the factor loadings to average 1 in each group (for each factor) and the intercepts to sum to 0 in each groups (for each factor again)Everything else is freely estimated.
Is it possible to implement this in Mplus (I believe so) ? If it is, how would you implement it in Mplus ?
Thanks a lot !
Linda K. Muthen
posted on Friday, May 16, 2008 - 11:11 am
You should be able to do this using MODEL CONSTRAINT.
Alexandre Morin
posted on Friday, May 16, 2008 - 1:28 pm
Hi again,
Thats what I thought. But I'm not overly familiar with this Mplus command (always got me a bit confused on how to use it...), could you give me a little push ?
Thanks in advance.
Linda K. Muthen
posted on Friday, May 16, 2008 - 2:15 pm
Why don't you give it a try and if you fail send your input, data, output, and license number to
Alexandre Morin
posted on Monday, May 19, 2008 - 3:38 am
Got it! thanks!
More simple than I thought (If I'm right - if not correct me). For a single group, that would give:
f1 BY y1* (c1)
y2 (c2)
y3 (c3);
f2 BY y4* (c4)
y5 (c5)
y6 (c6);
[y1] (c7);
[y2] (c8);
[y3] (c9);
[y4] (c10);
[y5] (c11);
[y6] (c12);
c1 = 3 - c2 - c3;
c4 = 3 - c5 - c6;
c7 = 0 - c8 - c9;
c10 = 0 - c11 - c12;
Linda K. Muthen
posted on Monday, May 19, 2008 - 8:10 am
The best way to tell if an input produces what you want is to run it and look at the results.
Alexandre Morin
posted on Tuesday, May 20, 2008 - 12:24 am
Yes, sorry I did not specify it. It does run correctly and provide fit indices equal to those obtianed under different constraints (marker variables, latent standardization). I was just wondering if t could be simplified. For instance, I tried
f1 BY y1* y2 y3 (c1-c3);
and it did not work (told me I had more constraints than variables).
But this way, everything is alright.
Thanks again.
Linda K. Muthen
posted on Tuesday, May 20, 2008 - 6:45 am
You can use a list of labels only with a list of variables. This is why it did not work.
Alexandre Morin
posted on Tuesday, May 20, 2008 - 7:48 am
Thank you very much Linda!
For those who followed this discussion, the previous input can thus be simplified to (and it work):
f1 BY y1* (c1)
y2-y3 (c2-c3) ;
f2 BY y4* (c4)
y5-y6 (c5-c6);
[y1-y6] (c7-c12);
c1 = 3 - c2 - c3;
c4 = 3 - c5 - c6;
c7 = 0 - c8 - c9;
c10 = 0 - c11 - c12;
guillermo montt
posted on Thursday, July 17, 2008 - 2:21 pm
I have recently done some multi-group confirmatory factor analysis models to test for measurement invariance across three groups. I tested for factor loadings, intercepts, and residual variances, but while invariance held for the first two, it did not for residual variances.
I know that measurement invariance requires the three to hold, but what does the above mean in terms of the interpretability of the estimates? Because the intercept and the loadings are equal, can the estimates be compared across groups? Is it just that the precision is different? What limits does this pose on comparative analyses?
Linda K. Muthen
posted on Friday, July 18, 2008 - 8:40 am
We are not of the school that residual variances need to be invariant.
Yara Hahr Marques H�kerberg
posted on Monday, August 10, 2009 - 11:37 am
Dear Mr and/or Ms Muth�n,
I am having a problem checking for measurement invariance of Demand Control Questionnaire in hospital workers of Brazil and Sweden using multiple group analysis. When I performed Confirmatory Factor Analysis for each country separately, using WLSMV estimator for categorical variables, I found that the best fit model was with 3 factors:
D1 by i1�i5, D2 by i6-i8 and D3 by i9-i10) for both countries, but with different crossloadings (Brazil: D1-i6 and Sweden: D1-i8). Is it possible to proceed with multiple group analysis? Do these models have equal factorial structure?
When I tried multiple group analysis, not considering the crossloadings, first of all, I fixed the highest loadings of each dimension in 1. After that, I used the default of Mplus (the loading of the first item of each dimension). However, using this procedure I couldn�t check equal loadings for the items I have fixed in 1, so I repeated the procedure fixing each factor variance in 1, but the results were totally different.
Was it correct? What procedure should I use? And why do the results differ?
Thanks in advance,
Linda K. Muthen
posted on Tuesday, August 11, 2009 - 11:13 am
If you have one factor loading fixed to one and the factor variance free or the factor variance fixed to one and all factor loadings free, you should get the same chi-square value. If you do not, please send your full outputs and license number to
Yara Hahr Marques H�kerberg
posted on Thursday, August 13, 2009 - 7:01 am
Thanks, Linda.
The chi-square value are very similar, but not equal. I will send you the outputs.
Is it correct to fix the factor variance in 1 to check equal loadings?
And, how about the factorial structure? Do you think I should proceed?
Thanks again,
Pranav Gandhi
posted on Tuesday, October 27, 2009 - 7:18 pm
Is it possible to assess measurement invariance over two time points controlling for covariates (e.g. comorbidities)? Could you suggest to me any references for this?
Linda K. Muthen
posted on Wednesday, October 28, 2009 - 6:22 am
I don't know of a reference but we show how to do this in our Topic 4 handout in the example for multiple indicator growth.
Pranav Gandhi
posted on Wednesday, October 28, 2009 - 8:54 am
Thank you for your reply.
But, will it be possible to estimate response shift using Oort or Schmitt's approach when controlling for covariates (e.g. comorbidities)? Also, I have data for two time points.
Linda K. Muthen
posted on Wednesday, October 28, 2009 - 9:48 am
I am not familiar with either Oort or Schmitt's approach. You can test for measurement invariance across time for two or more timepoints using the inputs shown in the Topic 4 handout.
Mario Mueller
posted on Tuesday, July 06, 2010 - 9:32 am
When intercepts are fixed by default how can I free them in longitudinal CFA (6 times/ 4 indicators)? When I do as recommended in the manual I get the error message:
... and problems with the alpha-parameter.
Linda K. Muthen
posted on Tuesday, July 06, 2010 - 9:55 am
Intercepts are not held equal across time as the default. Are you treating each time point as a group in a multiple group analysis?
Mario Mueller
posted on Wednesday, July 07, 2010 - 4:28 am
Thank you for your response...
Do you recommend to treat time points as multiple groups? This is my unrestricted model which has 6 latent variables (times 1 to 6)
model: sns1 by lxscl07 lxscl16 lxscl35 lxscl62 ;
sns2 by lzscl07 lzscl16 lzscl35 lzscl62 ;
sns3 by lqscl07 lqscl16 lqscl35 lqscl62 ;
sns4 by lvscl07 lvscl16 lvscl35 lvscl62 ;
sns5 by lwscl07 lwscl16 lwscl35 lwscl62 ;
sns6 by lbscl07 lbscl16 lbscl35 lbscl62 ;
in a next model I added residual covariances, then tested altered items to be fixed. Then I constrained factor loadings to be invariant over time. But for a test of strong factorial invariance I have to fix the intercepts to be invariant over time...How can I do this?
Mario Mueller
posted on Wednesday, July 07, 2010 - 5:25 am
I think I got it:
[lxscl07 lzscl07 lqscl07 lvscl07 lwscl07 lbscl07] (5);
[lxscl16 lzscl16 lqscl16 lvscl16 lwscl16 lbscl16] (6);
[lxscl35 lzscl35 lqscl35 lvscl35 lwscl35 lbscl35] (7);
[lxscl62 lzscl62 lqscl62 lvscl62 lwscl62 lbscl62] (8);
1 to 4 were the factor loading constraints.
Thanks anyway!
Linda K. Muthen
posted on Wednesday, July 07, 2010 - 6:08 am
Time points should not be treated as groups. They are not independent. Testing for measurement invariance across time is shown in the Topic 4 course handout starting with slide 78.
Richard E. Zinbarg
posted on Wednesday, October 06, 2010 - 9:16 pm
I am running into a puzzling result when trying to compare configural versus metric invariance of factor loadings across groups. I have a model that is essentially configurally invariant (I had to constrain a few loadings to be equal across groups to prevent some Heywood cases) and am comparing it with a metric invariant model in which all the loadings are constrained to be equal across groups. My understanding was the configural invariant model could not have a larger chi-square than the metric invariant model but this is precisely the result I am getting. Is my undestanding incorrect or does this seem odd to you too?
Rick Zinbarg
Linda K. Muthen
posted on Thursday, October 07, 2010 - 6:29 am
The more restrictive model should have the higher chi-square. It would be impossible to say more without seeing both outputs and your license number at
Gregory Arief D. Liem
posted on Sunday, May 15, 2011 - 11:01 pm
Hi Linda,
I am doing a multigroup invariance test. When I constrained parameters (factor loadings or residual variances or factor correlations or all three together), only the unstandardized parameters are constrained (having equal values for the two groups), whereas standardized parameters remain different in values Am I doing the right thing?
Really appreciate your advice.
Linda K. Muthen
posted on Monday, May 16, 2011 - 5:57 am
The standardized coefficients will be different even when the unstandardized coefficients are equal because the standardization is done using the standard deviations for each group not the overall standard deviations.
Vandita Vasudevan
posted on Wednesday, June 22, 2011 - 12:48 pm
Hi Dr. Muthen,
I am working on cross validating a Second Order CFA with MPlus. I was wondering if you would be able to guide me through this process.
Do you have any suggestions on the preferred order for imposing constraints on a 2nd order CFA when you start with a fully non-constrained model (intercepts & loadings free across groups � but means & intercepts of latents set to zero for identification purposes).
Thank you so much for your help,
Linda K. Muthen
posted on Wednesday, June 22, 2011 - 3:48 pm
I would test the first order factors first using the strategy shown in the Topic 1 course handout on the website. See multiple group analysis. Once measurement invariance is established for the first-order factors, I would test if for the second-order factor.
Vandita Vasudevan
posted on Wednesday, June 22, 2011 - 4:12 pm
I have tested for measurement variance among the first order factors. They cross validated well now I am planning to do the second order CFA. Are the steps in measurement invariance similar to that in first order CFA? I believe the means of the first order should be set to 0, is that right?
Step 1- fully non-constrained model
Step 2- constrain factor loadings
step 3- constrain intercepts and loadings
step 4- constrain intercepts, loadings, and residual variances
step 5- constrain intercepts, loadings, and residual variances and error variances
step 6- constrain intercepts, loadings, and residual variances, error variances, and covariances
Do these look correct?
Linda K. Muthen
posted on Thursday, June 23, 2011 - 11:06 am
Kristine Khachatryan
posted on Saturday, September 10, 2011 - 9:23 am
Dear Prof. Muthen,
I am running MGCFA with a four factor-model where two factors have only one indicator. I have already checked for the metric equivalence and got an acceptable model fit. In the next step by checking for the scalar invariance, the modification indices suggest to free the intercepts of two indicators (y8, y9), both loading on the same factor. If I do so I get this error message �THE STANDARD ERRORS OF THE MODEL PARAMETER ESTIMATES COULD NOT BE COMPUTED. THE MODEL MAY NOT BE IDENTIFIED. CHECK YOUR MODEL. PROBLEM INVOLVING PARAMETER 48�. The parameter 48 is the alpha-value for my latent construct perform. Would you have any suggestions how I can solve this problem?
Here is my syntax:
trust BY y1 y2 y3 y4 y5;
y2 with y3;
y1 with y2;
diffreg BY y6;
gemidnt BY y7;
perform BY y8 y9;
Model west: [y8 y9];
Thank you in advance!
Bengt O. Muthen
posted on Saturday, September 10, 2011 - 9:54 am
You cannot have the measurement intercepts for all (both) your factor indicators be free and also the factor mean. Fix the factor mean at zero for perform in the west group.
Kristine Khachatryan
posted on Saturday, September 10, 2011 - 2:14 pm
thank you very much for you helpful response!
Lisa M. Yarnell
posted on Friday, January 06, 2012 - 9:53 pm
Hi, I have a multigroup CFA model with UVI identification. I first estimated loadings of items freely in each group, then constrained them to be the same across the two groups, to test for support for measurement invariance. So, in the STANDARDIZED solution for the constrained model, all items have the same loading except for the first item for each factor, which shows up as 1.00 in the first group, with no significance associated with it; but as a real estimate with a significance associated with it for the second group.
My question is: Why is the first item per factor set at 1.000 for the first group in the STANDARDIZED solution for the constrained model, especially if I used UVI and not ULI identification? For example, the standardized solution for the constrained model shows the following.
For boys:
For girls:
Lisa M. Yarnell
posted on Friday, January 06, 2012 - 10:04 pm
Actually, I think I figured out the answer to my own question, so you can ignore the previous post! I think the answer is that even when constraining the loadings to be the same across the groups, I still need to have a * (an asterisk) for the first loading for each factor, in order for UVI and not ULI identification to be used!
posted on Wednesday, March 21, 2012 - 9:30 am
I am testing measurement invariance of a measurement model across two different age groups. However, when I restrict the factor loadings to be invariant across groups using stratified, weighted, and clustered data (WLSMV estimator) with categorical ordinal response scales, I receive the following error:
In addition, I noticed that I obtain somewhat different fit index estimates depending on whether I use Mplus 5.2 or 6.1. Any reason?
Linda K. Muthen
posted on Wednesday, March 21, 2012 - 10:54 am
The difference between the fit indices is because in Version 6 a new method for second-order chi-square adjustment for WLSMV, ULSMV, and MLMV resulting in the usual degrees of freedom was introduced.
Regarding the other problem, please send your output and license number to
Heather Whitcomb-Starnes
posted on Thursday, March 22, 2012 - 12:46 pm
I am trying to test for partial scalar invariance by freeing the intercepts on the items in 3 of my 6 factors, but I am getting the following error
Here are my model statements
f_infr by
parkdcar grass_strip streetlights;
f_conn by
altroute strght_st;
f_aest by
intrstng_thg natrlsight attrctv_bldg;
f_traf by
trffcspd_slow trffcspd_fast;
f_pers by
walk_unsafed walk_unsafen;
f_acce by
walkplaces walktobus;
MODEL Massachusetts:
[sidewalk parkdcar grass_strip streetlights
walkstores walkplaces walktobus
traffic trffcspd_slow trffcspd_fast]; !intercepts allowed to vary
Do I need to do fix the latent means @0 for these three factors in the Massachusetts groups?
Linda K. Muthen
posted on Thursday, March 22, 2012 - 12:48 pm
When intercepts are free, factor means must be fixed at zero. See the Topic 1 course handout under Multiple Group. All of the inputs for testing measurement invariance are given.
Hans Leto
posted on Monday, April 23, 2012 - 11:44 am
Dr. Muthen.
I am having problems testing measurement invariance with 2nd order factors. I am following the procedure described in the handout number 1. I do not know how to include a 2nd order factor in the example described in the slide 210.
Could you provide me more guidance. I describe an example (F3 is the 2nd order factor):
Model: f1 by y1-y5;
f2 by y6-y10;
F3 by f1 f2;
Model g2: f1 by y2-y5;
f2 by y7-y10;
F3 by f1 f2;
This syntax gives me an error.
Thank you in advance.
Linda K. Muthen
posted on Monday, April 23, 2012 - 12:31 pm
I suspect the problem is that a second-order factor is not identified unless it has three or more first-order factors.
Hans Leto
posted on Monday, April 23, 2012 - 12:51 pm
Thank you but it was just an example, the real one has more than three factors. Is the syntax correct? fixing to 0 the first order factors in both groups? It is just like the one showed in the slide.
Hans Leto
posted on Monday, April 23, 2012 - 12:58 pm
This will be more like the real one.
Model: f1 by y1-y5;
f2 by y6-y10;
f3 by y11-y15;
f4 by y16-20;
F5 by f1 f2 f3; �2nd order just by f1f2f3
] !all 1st order fact. fixed to 0
Model g2: f1 by y2-y5;
f2 by y7-y10;
f3 by y12-y15;
f4 by y17-20;
F5 by f1 f2 f3;
] !just 1st factor of the 2nd order fixed to 0
Bengt O. Muthen
posted on Monday, April 23, 2012 - 6:52 pm
It isn't clear if you get a syntax error or a modeling error. I'll address both.
You may get a syntax error by your statements
] !just 1st factor of the 2nd order fixed to 0
because you don't end them with semi colons. On the other hand, what you are posting may not be what you use in your run.
You will get a non-identification error because your second group frees up the intercepts for your y's which means that the factor mean difference in the second-order factor cannot be identified. Leave out the statement
and the default will give you the correct equality across groups of these intercepts.
Hans Leto
posted on Tuesday, April 24, 2012 - 2:33 am
Thank you for your answer. But it does not work, is not a syntax error (sorry). It is an error about "THE STANDARD ERRORS OF THE MODEL PARAMETER ESTIMATES COULD NOT BE COMPUTED."
I am quite new in testing invariance. My problem is specifying the 2nd order factors, because I already tested it only for 1st factors and ran perfectly.
I left out the [y1 y20], but it did not work.
My questions would be:
1. Do I have to fix to 0 the 2nd order factors in the general model or only the 1st order(in my example I only fixed to 0 the 1st order factors)?
2. In the specific group (g2)do I have to fix to 0 the 1st order factors (f1-f3) of my 2nd order?
3. Do I not have to free up the intercepts for the items in g2([y1-y20])?.
I have tried all these but still gives me the same error. I do not know what I am missing.
Thank you very much.
Bengt O. Muthen
posted on Tuesday, April 24, 2012 - 6:22 am
We can guide you better if you send your full output to Support.
J.D. Haltigan
posted on Monday, May 14, 2012 - 1:05 am
a quick question re: testing for configural invariance (equal form) that may have a simple explanation: when I test for equal form with both groups, my df and chi square do not equal the sum of the df and chi-square when I test each group separately (using USEOBSERVATIONS). As an example my df for both groups is 13, yet my df in the model for equal form = 31, not 26. Is this due to a mis-specification somewhere on my part? As an aside, rather than use a marker indicator I am fixing variances to 1 and I was wandering if this might make a difference though I doubt it. Group sizes are 131 and 128.
Linda K. Muthen
posted on Monday, May 14, 2012 - 5:38 am
Which estimator are you using and which version of Mplus?
J.D. Haltigan
posted on Monday, May 14, 2012 - 9:18 am
I am using Mplus v. 6.12 and estimator = ML.
Linda K. Muthen
posted on Monday, May 14, 2012 - 10:10 am
Please send the relevant outputs and your license number to
Sarah Phillips
posted on Friday, June 01, 2012 - 1:12 pm
I'm hoping to double check that I'm using the correct code for testing measurement invariance of some scales across different racial/ethnic groups. I've been using the video and handouts from Topic 1, but I'm trying to test invariance across 3 groups instead of 2. Would you mind confirming that I've got the correct code for the second test-- without invariance?
BE by m_a3 m_a31 m_a28 m_b50;
EE by m_a62;
CE by m_a72 m_a74 m_a79 m_a80;
Model AfAm:
BE by m_a31 m_a28 m_b50;
EE by ;
CE by m_a74 m_a79 m_a80;
Model Latino:
BE by m_a31 m_a28 m_b50;
EE by ;
CE by m_a74 m_a79 m_a80;
Thanks! -Sarah
Linda K. Muthen
posted on Friday, June 01, 2012 - 4:50 pm
I would remove EE BY in each group-specific MODEL command. With one indicator, you can't test for measurement invariance.
Sarah Phillips
posted on Monday, June 04, 2012 - 8:09 am
Thank you!
J.D. Haltigan
posted on Wednesday, July 18, 2012 - 10:59 am
I have a question concerning measurement invariance, specifically when testing for equal latent variances and latent means (i.e., population heterogeneity). If one chooses to freely estimate all indicators in baseline models (and ID'ing the model by setting the latent variances to 1), is it the case that one must use a separate baseline model in which marker indicators at 1 and variances are freely estimated if one wants to subsequently test for population heterogeneity (i.e., the baseline model must have variances freely estimated to subsequently test for equal latent variance(s))?
Thanks in advance!
Linda K. Muthen
posted on Wednesday, July 18, 2012 - 11:24 am
I think you are asking if it matters whether you set the metric of the factors by having a loading fixed at one or the factor variance fixed at one when you later compare structural parameters. If you fix the factor variances to one, you need to do this in only one group so the test of whether the factor variances are different across groups is a test of factor variances one in one group and free in the others verus factor variances one in all groups.
J.D. Haltigan
posted on Thursday, July 19, 2012 - 1:11 pm
Thanks Dr. Muthen.
I will try to articulate my question more clearly:
In my case I have a 2-factor model. I identify the model by setting the factor variances to 1 rather than the marker indicators for both groups. In other words, in the equal form solution my variances are already fixed to 1 in both groups (to ID the model) so no meaningful comparison could be made via the chi-square test of model fit for the subsequent test of invariant factor variances across groups, right? So to test for factor invariance, I would use a baseline model instead where I ID'd the model by using marker indicators rather than variances?
Linda K. Muthen
posted on Thursday, July 19, 2012 - 3:45 pm
In multiple group analysis, you need to fix the factor variances to one in only one group. They can be free in the other groups. A meaningful test of whether the variances differ across groups is a test of factor variances one in one group and free in the others verus factor variances one in all groups.
Aaaron Chan
posted on Friday, July 20, 2012 - 11:36 pm
Hi, I am working on a multi-group CFA for testing measurement invariance across 5 samples. The hypothesized model is a two order factor model. Aiming at testing metric invariance, the following syntax failed to work. What is the problem?
Thank you for your advice.
DATA: FILE IS data.prn;
VARIABLE: NAMES ARE e1-e3, m1-m4,d4-d5,d9, g;
USEVARIABLE ARE e1-e3, m1-m4,d4-d5,d9, g;
GROUPING IS g (1=g1, 2=g2, 3=g3, 4=g4, 5=g5);
D BY d4*, d5, d9;
; [
E BY e1*, e2-e3;
; [
M BY m1*, m2-m4;
; [
Model g1:
[d9]; !allow the intercepts to differ
Model g2:
[d9]; !allow the intercepts to differ
Model g3:
[d9]; !allow the intercepts to differ
Model g4:
[d9]; !allow the intercepts to differ
Model g5:
[d9]; !allow the intercepts to differ
Linda K. Muthen
posted on Saturday, July 21, 2012 - 9:51 am
When the intercepts are free, all factors means must be zero. The mean of g is not fixed at zero.
Fred Danner
posted on Friday, February 15, 2013 - 11:33 am
Hi, I am testing second-order measurement invariance, using MLR estimation. Unconstrained model gives reasonable results. Model constraining factor loadings runs fine but cuts the N in each group in half! Why??
f1 by x1 - x6;
f2 by x7 - x10;
f3 by x11 - x13;
f4 by f1 f2 f3;
[f1 - f3 @0];
Model g2:
f1 by x2 - x6;
f2 by x8 - x10;
f3 by x12 - x13;
[x1 - x13 f1 - f4 @0];
f1 by x1 - x6;
f2 by x7 - x10;
f3 by x11 - x13;
f4 by f1 f2 f3;
[f1 - f3 @0];
Model g2:
[x1 - x13 f1 - f4 @0];
Bengt O. Muthen
posted on Friday, February 15, 2013 - 1:33 pm
We have to see the 2 outputs to be able to tell. Please send to Support.
Fred Danner
posted on Friday, February 15, 2013 - 1:57 pm
I have done so, but please do not waste your time poring over it -- I am embarrassed to say that I inadvertently listed one extra variable on my input list. Removing this variable fixed the problem.
Elizabeth Shulman
posted on Friday, March 01, 2013 - 9:06 am
Dear Drs. Muthen,
I have two questions about using TYPE=CLUSTER in data sets that have repeated observations of the same individuals.
1) In one data set, I have measures at two time points, six years apart. I am entering data into MPLUS in the long format and specifying my DV as a latent variable, which is regressed on age. I use TYPE=COMPLEX and cluster on subject ID. Is there a name for this sort of analysis?
2) In another data set with 1 to 7 repeated measures of the same individuals, I wanted to compare age groups' (adolescent vs. adult) means on a given variable, even though they are the same individuals. I ran a simple regression with the age groups entered as dummy variables. Again, I imported the data in long format and used TYPE=CLUSTER to cluster on subject ID. Is there any reason that it would be incorrect to draw inferences about the mean differences between the age groups based on this regression?
Elizabeth Shulman
posted on Friday, March 01, 2013 - 10:40 am
P.S. I did not mean to write "TYPE=CLUSTER" in the first sentence of the above post, but rather TYPE=COMPLEX (in conjunction with a CLUSTER command).
Linda K. Muthen
posted on Friday, March 01, 2013 - 6:15 pm
1. I know of no special name for this model. It is a latent variable model.
2. This sounds okay.
Tom Booth
posted on Saturday, March 09, 2013 - 10:42 am
I am trying to fit a second-order invariance model with categorical indicators using the delta method for 2 groups. I was interested in following the suggestion of Chen, Sousa and West (2005) and testing invariance in the following order;
1: Configural
2: 1st order metric (loadings)
3: 2nd order metric (loadings)
4: 1st order scalar (thresholds)
5: 2nd order scalar (intercepts)
Where the following constraints are used across groups in each model:
1: First and second order loadings free in both groups (first item/factor loadings fixed to identify). Item thresholds free in both groups. First and second order factor means fixed to 0 in both groups. Scale factors fixed at 1 in both groups.
2: As (1) but with first order loadings constrained equal.
3: As (2) but with second order loadings constrained equal.
4: As (3) but with item thresholds constrained equal, first order factor means free in group 2, and scale factors free in group 2.
5: As (4) but with second order factor mean free in group 2 and first order factor means constrained equal.
I am not sure if this sequence is correct and after noting discussion here and notes on the Mplus site on the Millsap and Tien (2004) paper, I fear I have missed something crucial. Any guidance on the matter would be much appreciated.
Bengt O. Muthen
posted on Saturday, March 09, 2013 - 11:29 am
There are different approaches for binary and polytomous items. With binary items, Step 2 adds scale factors differences across groups which makes the model not identified when the thresholds are different. With polytomous items, the Millsap-Tien approach can be followed.
Tom Booth
posted on Saturday, March 09, 2013 - 11:35 am
Hi Bengt,
Thank you for the very swift response. Just for clarity, my items are polytomous. From your response, I take it that in principal there is no issue following the Chen, Sousa and West sequence, so long as the identification constraints of Millsap-Tien are followed, and that these are different to the basic model specs I note above?
Bengt O. Muthen
posted on Saturday, March 09, 2013 - 11:52 am
Yes, the steps you list look fine.
Tom Booth
posted on Sunday, March 10, 2013 - 3:35 am
Thanks Bengt. I had thought from the discussions that with the categorical nature of the data and use of WLSMV, loadings and thresholds needed to be considered together, not split as in the above stages.
Tom Booth
posted on Sunday, March 10, 2013 - 4:52 am
Sorry, I have a further follow up question. Within the sequence of models above, when thresholds are constrained across groups, scale factors are freed in the second group. I have 3 questions on this;
1- Is this correct?
2- Is this necessary?
3- If one then subsequently releases thresholds, partial invariance, do the associated item scale factors need to be fixed again?
Bengt O. Muthen
posted on Sunday, March 10, 2013 - 1:00 pm
Re your 3:35 post:
Loadings and thresholds are considered together in the binary case.
Re your 4:52 post:
1. Scale factors are needed whenever you make comparisons of the factors, that is, in the metric and scalar cases.
2. Yes, because scale factors contain 3 things: Loadings, factor variances, and residual variances. So even when loadings are invariant, scale factors won't be - in particular you want to take into account the factor variance variation across groups.
3. You fix scale factors in the configural case because in that case you are not comparing factors across groups.
Tom Booth
posted on Sunday, March 10, 2013 - 1:49 pm
Thank you very much.
Michelle Little
posted on Friday, June 21, 2013 - 12:40 pm
Hello Linda and Bengt,
I am doing a 4 group test of measurement invariance with ordered categorical items (4-point response set). The measure is invariant on loadings, but not on thresholds.
I would like to examine specific contrasts (ethnicity within gender and gender within ethnicity). Do you know of any problems with using the MODEL CONSTRAINT command to simultaneously examine threshold differences across items per my contrasts of interest? I am thinking it may simplify the analyses. I could not find an example in mplus examples or the literature...
Thanks in advance for your help.
Michelle L.
Bengt O. Muthen
posted on Friday, June 21, 2013 - 3:43 pm
It makes senses to do such testing.
Leslie Rutkowski
posted on Wednesday, September 18, 2013 - 1:01 pm
I'm wondering how to specify a multiple-group model where some groups have a subset of the total items in a scale.
A simplified example results in errors:
With the errors:
Variable is uncorrelated with all other variables: V3
Group G1 does not contain all values of categorical variable: V3
Bengt O. Muthen
posted on Wednesday, September 18, 2013 - 1:52 pm
See our FAQ:
Different number of variables in different groups
Leslie Rutkowski
posted on Wednesday, September 18, 2013 - 2:01 pm
Fantastic, Bengt. Thank you!
Leslie Rutkowski
posted on Monday, September 23, 2013 - 10:23 am
Hi again. I have a follow up to the last question. I followed the FAQ and used
MODEL: f1 BY var1 var2 var3 var4 ;
var4 (2);
f1 BY var1 var2 var3 var4 ;
var4 (2);
This still produces the error that var4 has no non-missing values in group 2. Am I missing a step?
Bengt O. Muthen
posted on Monday, September 23, 2013 - 3:32 pm
Try VARIANCES = NOCHECK in the DATA command and if that doesn't resolve it, send files to Support.
posted on Tuesday, October 15, 2013 - 6:49 am
Dear Linda and Bengt,
My question is the following, which has been asked before:
I would like to test for measurement invariance using the difference in McDonald's non-centrality index (NCI) as recommended by Meade et al (2008) in "Power and Sensitivity of Alternative Fit Indices in Tests of Measurement Invariance" J Appl Psych.
You (Linda) replied that Mplus does not give an NCI index. However, since my sample is very big, I really would like to report it next to the CFI. Do you have any formula or idea how I can derive de NCI index (maybe from other given fit indices)?
Thank you in advance for your answer!
Kind regards,
Linda K. Muthen
posted on Tuesday, October 15, 2013 - 10:33 am
I would Google this to find the formula and then see if the information for computing it is available in the Mplus results. You may want to ask on a general discussion forum like SEMNET
posted on Wednesday, October 16, 2013 - 12:31 am
Thank you!
Kind regards,
posted on Tuesday, October 29, 2013 - 9:48 am
For all Mplus users who would like to know McDonald's NCI as well: I figured out how to calculate it (by hand) from the Mplus output.
The formula is as follows:
exp(-0.5*((Chisquare of the target model - DF of the target model)/(N-1)))
Where 'exp' = exponantial (you can calculate this by using for example this website
You calculate the formula twice: once for you Configural Invariance model and once for your Measurement Invariance model. Then, you substract the value of the CI model from the value of the MI model. This is the final Mc Donald's NCI difference value you can report. The cut-off value for an invariant model differs per number of factors and items. In the article of Meade and Johnson (2008) you can find at page 586 a table with these cut-off points. (Meade, AW & Johnson, C (2008). Power and sensitivity of alternative fit indices in tests of measurement invariance. Journal of applied psychology, 93, 568-592)
Linda K. Muthen
posted on Tuesday, October 29, 2013 - 10:05 am
Thank you so much for sharing that information.
Ian Koh
posted on Friday, December 13, 2013 - 12:26 am
Dear Bengt and Linda,
I ran a test for factorial invariance (six-factor structure, with partial measurement invariance) across two groups following the steps mentioned in Byrne (2011). Out of curiosity, I'd like to ask: Are the configural model's parameters estimated using the whole sample, or are they estimated from the group samples? Thanks for your help.
Linda K. Muthen
posted on Friday, December 13, 2013 - 7:06 am
The model for each group is estimated using the data from that group.
Ian Koh
posted on Monday, December 16, 2013 - 4:38 pm
Thanks Linda! This question follows from my previous post (dated Friday, 13 December 2013).
Before fitting the configural model, I first fitted two baseline models: one for 5-year-olds and one for 6-year-olds. The 5-year-old group didn't require any modifications to the original model specification; however, the 6-year-old group required one extra cross loading, else there would've been a nonpositive definite matrix message. My configural model converged without any issues when including the extra cross loading for the 6-year-old group (as expected). However, the configural model also converged without any nonpositive definite matrix message when the cross loading was removed.
I also tested for factorial invariance over gender using the same model specification, encountering the same issue for the gender baseline models. (Namely, that the female group required one extra cross loading so that its solution would not have a nonpositive definite matrix error, while the male group required no modifications.) What puzzles me is that this nonpositive definite matrix issue was replicated in the gender configural model simply by removing the extra cross loading for the female group, but specifying the cross loading resulted in an admissible solution.
Why do these two configural models behave differently?
Thanks again,
Bengt O. Muthen
posted on Tuesday, December 17, 2013 - 8:47 am
Please send outputs and data if possible. Let's focus on the 5 vs 6 year old runs, so send the 6-year old separate run with and without the cross-loading and the 2-group run of 5 and 6 year olds with and without the cross-loading.
Ellyn L.
posted on Thursday, February 06, 2014 - 12:25 pm
Drs. Muthen and Muthen,
I am conducting a multiple group analysis and need to asses mean invariance. I have written syntax that runs successfully, but I'm not sure that I'm including (all of) the correct code. I have consulted both the Mplus user guide and blog posts, and I am looking for some confirmation/input on the syntax I am using to assess mean invariance. I have included the Model input information below. Thanks so much.
I ON M So;
Su ON I P N;
Sh ON I Su So P N;
D ON Sh;
E ON Sh;
M WITH So P N;
So WITH P N;
Linda K. Muthen
posted on Thursday, February 06, 2014 - 1:26 pm
If you want to test if means are equal across groups, compare the model where they are free across groups to the model where they are held equal across groups.
In a conditional model, it is the intercepts that are model parameters not the means.
Mosi Ifatunji
posted on Thursday, February 13, 2014 - 6:40 am
Drs. Muthen and Muthen,
I am conducting a multiple group confirmatory factor analysis with three comparison groups. The observed variables are categorical. I am using the Theta parameterization. The focus of the analysis is to test for construct invariance between the three groups. I currently have the factor variance, factor loadings and thresholds set to be estimated and equal between all groups (varying within groups, constrained between groups). I would like to do the same for the residual variances. However, as you know, when I use the Theta parameterization, the residual variances for the omitted group are set to 1. This means that to estimate the sought model of construct invariance, I must set the residual variances in the two comparison groups to 1. I have done this.
My question is: When I set the comparison group residual variances to 1, the values for the "Est./S.E." for the residual variances for the two comparison groups reads "Infinity." Is this a problem? Is there a fix for this or a work around? The output contains no fatal error reports.
Linda K. Muthen
posted on Thursday, February 13, 2014 - 11:11 am
Please send your output and license number to
posted on Monday, June 02, 2014 - 10:28 am
Drs. Muthen and Muthen,
I am running a multi-group CFA on the nutrition self-efficacy scale and want to test the measurement invariance across 9 coutries. I have followed the handout/video for topic 1 and done the following syntax:
Usevariables are Q3r1 Q3r2 Q3r3 Q3r4 Q3r5 weight;
Grouping is Country (1 = Norway 2 = Germany 3 = Spain 4 = Greece
5 = Poland 6 = UK 7 = Irel 8 = NL 9 = Portugal);
weight is weight;
Analysis: estimator = MLR;
Model: NSE BY Q3r1-Q3r5;
Model Norway: NSE BY Q3r2-Q3r5;
standardized; modindices (3.84);
My question is how do I add in the other countries. Should I continue to add the syntax in for Model Germany, Spain etc the same as Model Norway?
Linda K. Muthen
posted on Monday, June 02, 2014 - 10:35 am
See the Version 7.1 addendum on the website with the user's guide. There are convenience features for testing measurement invariance across groups. You may find them helpful.
posted on Monday, June 02, 2014 - 10:43 am
I have saw the convenience features for 7.1, but current am working from version 7.
Linda K. Muthen
posted on Monday, June 02, 2014 - 11:44 am
You would do the same thing you have done for Norway for each group.
posted on Monday, June 30, 2014 - 9:45 am
Following on from my previous question, a collegue suggested this: Test for the equality of the loadings, still allowing item and factor intercepts to vary. Then test for the equality of the item intercepts, still allowing factor intercepts to vary. Finally test for the equality of the factor intercepts.
How do I allow factor intercepts to vary across countries, how would this look in syntax?
MODEL: SEffic BY
SEffic BY Q3r2* (p2);
SEffic BY Q3r3* (p3);
SEffic BY Q3r4* (p4);
SEffic BY Q3r5* (p5);
[Q3r1*] (p6);
[Q3r2*] (p7);
[Q3r3*] (p8);
[Q3r4*] (p9);
[Q3r5*] (p10);
Q3r1 WITH Q3r2* (p12);
MODEL 2: SEffic BY
SEffic BY Q3r2* (p2);
SEffic BY Q3r3* (p3);
SEffic BY Q3r4* (p4);
SEffic BY Q3r5* (p5);
[Q3r1*] (p6);
[Q3r2*] (p7);
[Q3r3*] (p8);
[Q3r4*] (p9);
[Q3r5*] (p10);
Q3r1 WITH Q3r2* (p12);
Many Thanks,
Linda K. Muthen
posted on Monday, June 30, 2014 - 10:21 am
See the Topic 1 course handout on the website where the input files for testing for measurement invariance are shown under multiple group analysis.
Amy Walzer
posted on Tuesday, August 26, 2014 - 7:35 pm
I have tested measurement invariance using the new convenience features in MPlus (i.e., ANALYSIS: MODEL=CONFIGURAL METRIC SCALAR) to see if there is measurement invariance in my measure between men and women.
Now, I'd like to go on to test other types of invariance outlined by Steinmetz et al., 2009:
1.) Invariance of error variance
2.) Invariance of factor variance
3.) Latent value means
4.) Factor covariance
How do I go about doing this?
When I try to build the syntax so that it constrains the necessary elements (e.g., error variances, the variance of each of the factors) in my second group (women) to equal my first group (men), I get an error stating "Model did not terminate normally. Refer to TECH9 output for more information."
Thanks much.
Linda K. Muthen
posted on Wednesday, August 27, 2014 - 6:11 am
Please send the output and your license number to
David Vachon
posted on Saturday, January 03, 2015 - 10:33 am
I am a novice Mplus user and I am trying to test for measurement invariance on a model with censored (from below) indicators. I have 12 child maltreatment indicators loading on 4 latent variables. The User Guide has a nice step-by-step description of measurement invariance procedures for continuous and categorical outcomes (pp. 484-486), but nothing for censored outcomes. I have been trying to base my models on these recommendations, but I do not know enough about Mplus or censored analysis to feel confident that I am on the right track. Could you outline the steps used to test for measurement invariance with censored outcomes (i.e., what should I fix or free at each step)? I have been using the WLSMV estimator.
Kind regards,
Bengt O. Muthen
posted on Saturday, January 03, 2015 - 5:21 pm
You would use the same approach as for continuous outcomes.
Hugo Cogo-Moreira
posted on Wednesday, February 18, 2015 - 3:39 am
Dear Muth�ns,
Considering a invariance testing for unidimensional model, the estimation terminated normally and the scalar against configural model returned p = .0793.
For the configural and scalar models X� p-values are higher 0.05 and RMSEA lower than 0.06; however, CFI and TLI were below than 0.95 reducing to 0.87 in the scalar level test.
I found that "... CFI will keep decreasing as a model becomes more restrictive" in invariance testing and the authors concluded that it would be not useful for such purpose (Hong et all., 2003 Educational and Psychological Measurement, Vol. 63 No. 4, August 2003).
Would be the same applied to TLI?
Best wishes,
Bengt O. Muthen
posted on Wednesday, February 18, 2015 - 8:08 am
You may want to ask this general question on SEMNET.
Eric Deemer
posted on Wednesday, February 18, 2015 - 11:16 am
I'm using the new convenience feature to test the invariance of a CFA model across 3 groups. How does one know which groups are being compared when there are more than 2 groups? Would I use the ALIGNMENT option here?
Also, with the convenience feature is it no longer necessary to calculate the chi-square difference value by hand? The
website says chi-square difference testing is carried out automatically with version 7.1, but I still get a warning in my output saying that the chi-square value cannot be used in the regular way.
Linda K. Muthen
posted on Wednesday, February 18, 2015 - 11:18 am
All 3 groups are being compared. There is no need for alignment.
This is a warning that is always printed. It does not apply to you in this case.
Herv� CACI
posted on Monday, March 16, 2015 - 4:32 am
Dear all,
I came across an error message while conducting measurement invariance testing. I was testing a bifactor model for variances invariance between two age groups. After successfully testing for uniqueness invariance, I discarded the following line the 2nd group:
Inatt*; Hyper*; Imp*; g*; ! 3 specific factors and 1 g-factor
This is the only modification I made. Any idea?
Parameter 73, Group YOUNG: G BY I10 (equality/label)
Where the 'YOUNG' group is the first group...
Thank you in advance.
Linda K. Muthen
posted on Monday, March 16, 2015 - 6:29 am
Please send the output and your license number to
Raffaele Zanoli
posted on Sunday, March 22, 2015 - 11:17 am
Hello again. I ma still, having some problems with Muli Group invariance testing.
I am testing a multi-country (6) model and all goes fine but in one country. There I have that 2 latent variables have a correlation higher than 1. That means they are seen as same in that specific country but not in all others.
Is there a way I can run a model with one of the LV is dropped JUST for that country (group)?
In alternative, I am not sure it is acceptable (and will work) to constrain the two LV to have a variance just below 1 (0.99) by e.g. F1 WITH
Bengt O. Muthen
posted on Monday, March 23, 2015 - 1:53 pm
Sounds like that group is better suited for one factor which means it has nothing invariant with the other groups so it should not be part of the multi-group analysis.
I would not constrain the factor covariance that way.
If other countries also have high factor correlations, you might consider ESEM or BSEM multiple-group analysis to allow cross-loadings which tend to reduce factor correlations (see our website).
Paula Vagos
posted on Monday, April 20, 2015 - 4:12 am
Hello all, I am following up on the discussion on how to calculate the NCI based on mplus output. I am trying to use the formula provided by Marlies, and would like to ask for help regarding the calculation of exp... What values am i supposed to enter in the website that was recommended?
I am sorry if did is not a question suitable for this mplus discussion, but I found no where else to ask...
would really appreciate any help!
Linda K. Muthen
posted on Monday, April 20, 2015 - 10:31 am
I think this question is more appropriate for a general discussion forum like SEMNET.
Tahir Wani
posted on Friday, August 07, 2015 - 3:12 am
I got the following MODEL FIT INFORMATION while testing invariance,
Invariance Testing
Number of Degrees of
Model Parameters Chi-square Freedom P-value
Configural 392 2604.345 1958 0.0000
Metric 356 2664.537 1994 0.0000
Scalar 320 2732.325 2030 0.0000
Degrees of
Models Compared Chi-square Freedom P-value
Metric against Configural 61.251 36 0.0054
Scalar against Configural 129.983 72 0.0000
Scalar against Metric 68.576 36 0.0009
Is it right to assume that there is no Invariance across groups in this case. In other words is it right to assume that when the p value is significant there is no invariance?
Bengt O. Muthen
posted on Friday, August 07, 2015 - 4:56 pm
Yes. Although, more specifically, there is not full invariance. There could still be partial invariance (for all but a few items).
milan lee
posted on Friday, August 07, 2015 - 8:15 pm
Hello Dr. Muthen,
When testing the partial measurement invariance of factor loadings across three groups, I found that it was achieved after I freed up some loadings for each of the three groups. Specifically, 3 indicators were relaxed in the first group, 2 in the second group and 3 in the third group. Moreover, these freed-up indicators are not identical across groups. Can I still say that the partial metric invariance holds? Thank you!
Bengt O. Muthen
posted on Saturday, August 08, 2015 - 11:07 am
It is equalities across groups in say loading parameters that you should be relaxing. So I don't understand your statement that the they are "not identical across groups". Unless you mean that some parameters are unequal across 2 of the 3 groups and some other unequal across 2 other groups.
Tahir Wani
posted on Saturday, August 08, 2015 - 11:17 am
Dr Muthen,
After the invariance measurement I want to do a multi group analysis specifically find out group differences across all paths, does MPlus provide with a mean difference with any specific input command or does it need to be computed manually by chi square difference.
milan lee
posted on Saturday, August 08, 2015 - 1:49 pm
Thank you, Dr. Muthen!
Yes, I meant that.
So my questionnaire has 15 items and the partial metric invariance across 3 groups held when:
4 items/indicators were freed only in group2 (resulting in their loadings in this group to be different from the other two groups) and another 2 items were freed only in group3.
Most relevant literature considers that partial invariance holds as long as loadings of 2 items remain equivalent. But are there any criteria to rule partial metric invariance under the circumstance I encountered as above?
Again, thank you very much!
Bengt O. Muthen
posted on Sunday, August 09, 2015 - 7:22 pm
Answer to milan lee:
The statistical rule is that you want an identified model; the output will tell you if it isn't. The substantive rule is that you want to have sufficiently many equalities that you believe you are measuring the same thing.
Bengt O. Muthen
posted on Sunday, August 09, 2015 - 7:24 pm
Answer to Tahir Wani:
You need to test it yourself; either by chi-2 diffs or by Model Test.
Manuel Torres Sahli
posted on Thursday, August 20, 2015 - 9:07 am
I'm testing measurement invariance across gender groups, and statistical fit (CFI, RSMEA) is better in constrained models. I haven't seen that before, so I don't know if it's ok. If the difference (+) is bigger than cutoff criteria, should I worry? Or is ok if the constrained tests fit better, no matter how much.
Thank you in advance,
Linda K. Muthen
posted on Thursday, August 20, 2015 - 2:28 pm
Please send the outputs and your license number to
posted on Monday, September 28, 2015 - 4:51 am
A warning appears in my output file: "MODINDICES option is not available when performing measurement invariance
testing with multiple models with the MODEL option of the ANALYSIS command". I don't understand it because I've used modindices command performing measurement invariance and it worked.
My inp is:
Names are
v6 v394 v395 v396 v397 v398 v399;
usevariables are
Grouping is V6
(2=Belgium 3=Netherlands 4=Germany_West 5=Italy
model = configural metric scalar;
FOR by V396 V397 V398;
AG by V394* V395 V399;
stand TECH9; modindices;
Linda K. Muthen
posted on Monday, September 28, 2015 - 7:14 am
They are not available when you test for measurement invariance using the following options:
model = configural metric scalar;
Tahir Wani
posted on Monday, October 12, 2015 - 6:26 am
Dear Dr Muthen,
I found that my dataset was not normal so used MLR estimator. I check the model for invariance and found it invariant at configular, metric and scalar levels for all the 5 demographic variables I had used. Now I ran the model e.g for Gender(male and female) and got the measurement and structural paths for both males and females. I am now confused that should I report these paths and show how they differ for the other group or do I have to do chi-square difference test for the paths. If so can we do that for MLR and please if you can guide how to do it. I am familiar with ML and AMOS to perform it but since I am new to MPlus have got no clue.
Linda K. Muthen
posted on Monday, October 12, 2015 - 10:23 am
If you want to compare the paths across groups, you can do chi-square difference testing or use the Wald test of MODEL TEST. How to do difference testing using MLR is described on the webiste. See How To in the left column.
Masa Vidmar
posted on Monday, October 12, 2015 - 2:12 pm
I am running a multiple-group CFA with two factors. I did a weak invariance test across two groups (langugaes). I did not find support for the weak invariance (only configural).
Now I would like to test partial invariance by only constraining one loading at the time - I would like to examine if all indicators are a problem or possibly just one. Is that ok to do?
Also, I do not know how to write an input for this (for only constraining one loading). Can you help me? Below is the extract from the input for testing weak invariance:
model SLO:
pips_lit by Writing @1;
by iar (1)
by letters_p (2)
word read_p (3);
pips_mat by
by numbers (4);
by math (5);
[Writing] (100);
[word] ;
[sums] (200);
[numbers] ;
model GER:
pips_lit by Writing @1;
by iar (10)
by letters_p (20)
word read_p (30);
pips_mat by
by numbers (40);
by math (50);
[Writing] (100);
[word] ;
[sums] (200);
[numbers] ;
Bengt O. Muthen
posted on Monday, October 12, 2015 - 5:57 pm
To add an invariant loading, say e.g.
BY iar (1);
in Model GER.
Tahir Wani
posted on Tuesday, October 13, 2015 - 3:31 am
Dear Sir, Yes exactly I want to compare the paths across groups, but the thing is that I haven`t hypothesised to check the difference on any particular path, in other words I want to perform it on all the structural paths available. So I am not sure what a constrained or nested model here means and which one will be a baseline model. If you can please explain or help me with this matter.
Linda K. Muthen
posted on Tuesday, October 13, 2015 - 6:27 am
You can have all paths free and then constrain one to be equal at a time. Or you constraint them all to be equal. This choice must be yours.
Tais S. Barreto
posted on Monday, October 19, 2015 - 7:25 pm
I am trying to test measurement invariance across several cultural groups (about 11 countries). I watched the topic 1 video and see that most of the discussion is centered on testing invariance for two groups (male/female). Are there any issues with testing invariance on several groups and if not, do you recommend using MIMIC or multi-group CFA?
Thank you.
Bengt O. Muthen
posted on Tuesday, October 20, 2015 - 9:12 am
Several methods for many groups are shown at our Measurement Invariance page:
Jinni Su
posted on Friday, October 30, 2015 - 4:08 pm
Dear Dr. Muthen,
My colleagues and I ran multigroup CFA models to evaluate measurement invariance across race (European American vs African American). And reviewers commented that we may have established invariance across both race and income/SES given the possible confound between race and income/SES. Do you have any advice on how to tackle this issue? Is there a way to evaluate measurement invariance across race while controlling for SES?
Bengt O. Muthen
posted on Friday, October 30, 2015 - 4:50 pm
You may want to raise this question on SEMNET.
posted on Wednesday, February 03, 2016 - 12:17 pm
I have a few of questions about CFA and measurement invariance across 11 groups. First, I run one-factor model with four nominal indicators 11 times, but the output didn't have the values of CFI, TLE,RMSEA. Is it because of nominal indicators?
Second question is that how can I do measurement invariance for nominal variables? I check the MODEL = CONFIGURAL METRIC SCALAR, however, it can't work for the nominal variable.
Bengt O. Muthen
posted on Wednesday, February 03, 2016 - 5:46 pm
Q1. Yes. Try TECH10.
Q2. You would have to specify invariance yourself.
posted on Thursday, February 04, 2016 - 6:48 am
I have tried "TECH 10" in the output command, but it still not showed CFI,TLE,and RMSEA.
As for testing measurement invariance on nominal indicators, I am not sure if my codes are specified correctly in the model command. Also, I run the codes, but it showed me"*** ERROR
Group AUS has 0 observations.
Group CAN has 0 observations.
Group GBR has 0 observations."
VARIABLE:NAMES ARE CNT u1-u4;
USEVARIABLES ARE u1-u4;
NOMINAL are u1-u4;
f1 BY u1-u4 ;
f1 BY u1-u4 ;
f1 BY u1-u4 ;
f1 BY u1-u4 ;
f1 BY u1-u4 ;
f1 BY u1-u4 ;
f1 BY u1-u4 ;
f1 BY u1-u4 ;
f1 BY u1-u4 ;
f1 BY u1-u4 ;
f1 BY u1-u4 ;
Linda K. Muthen
posted on Thursday, February 04, 2016 - 10:08 am
Please send the output, data set, and your license number to
Anders Hofverberg
posted on Friday, March 11, 2016 - 12:28 am
I am conducting a 2-group measurement invariance (MI) testing on a model with three latent factors and 13 continuous indicators. I am using MLR as estimator because of non-normality. I have used the MODEL IS CONFIGURAL METRIC SCALAR command and at least partial scalar MI is supported. Now I would like to test strict MI. Hence these questions:
1. Could strict MI be tested using MODEL IS SCALAR command and add restrictions for invariant variance across groups?
2. If yes, do I just add e.g. y1(1);y2(2)... under MODEL to keep variances equal across groups?
3. Is the Chi-square test provided with the MODEL IS CONFIGURAL METRIC SCALAR command already corrected for the MLR estimator? Or do I calculate the SB scaled chi-square by hand? My interpretation of the user guide is that the scaling correction is carried out automatically.
Thank you!
Linda K. Muthen
posted on Friday, March 11, 2016 - 7:28 am
1. I do not think so but you can try.
2. Yes. These are residual variances in the factor model.
3. Yes.
Paula Vagos
posted on Wednesday, March 30, 2016 - 2:41 am
Dear Doctor Muthens,
I am testing measurement invariance of a self-report instrument that uses a five-point likert type scale. The data is not multivariate normal, so I am using the MLR estimator. I tested configural, then metric and then scalar invariance, but I got the feedback that because I am using ordinal variables, I should only test for configural and scalar invariance.
I tried reading about this (for example, Milsap, 2004 at
) but still find it confusing.
I was wondering if you had an opinion on this and/or could point me to some relevant readings on this.
Thank you!
Bengt O. Muthen
posted on Wednesday, March 30, 2016 - 10:10 am
It is a bit confusing. Millsap's book on measurement invariance describes this. To quote an email from him: "The invariance constraints on the thresholds in my book on 5.19 are meant to apply to the configural case. Once metric invariance is imposed, one can actually release some constraints on the thresholds and still achieve identification." For this reason it may be safer to only test for configural and scalar invariance.
Paula Vagos
posted on Thursday, March 31, 2016 - 2:40 am
Thank Doctor Bengt,
It is actually very confusing for me...
I would dare ask you a follow up question. What I should do, then, is:
1. Test for configural invariance with all thresholds constraint to be equal and loadings free
2. Test for scalar invariance with all loadings and thresholds constraint to be equal
3. Free one threshold at a time in trying to achieve a non-significant chi-square difference?
Again, thank you for any help you might give me on this subject!
Bengt O. Muthen
posted on Thursday, March 31, 2016 - 3:18 pm
I would use our automatic Model = configural scalar approaches discussed in our UG. For scalar you can then go back and free parameters that have large Modindices.
Paula Vagos
posted on Friday, April 01, 2016 - 6:23 am
Thank you again Dr. Bengt,
I wasn't aware of this update.
Still, I had understood that Millsap had suggested constraining the thresholds/ intercepts when testing for configural invariance, whereas configural invariance as calculated using the MODEL = configural scalar option does not apply this constraint.
So, if I am understanding correctly, you are suggesting comparing the completely free model (i.e., configural) with the scalar model, disregarding an eventual threshold constraint at the configural level?
Thank you for your attention and help!
Bengt O. Muthen
posted on Friday, April 01, 2016 - 12:12 pm
Pia H.
posted on Monday, April 04, 2016 - 6:52 am
I just did invariance measurement using the MODEL = configural scalar;
PARAMETERIZATION = theta;
commands (I cannot check for metric MI because I have variables loading onto more than one factor).
The output gives me non-significant comparison for scalar against configural, which should be good.
But: the CFI values slightly improve from the configural (.942) to the scalar (.943) model. Are they not supposed to detoriate with adding more constraints?
Thank you very much,
Linda K. Muthen
posted on Monday, April 04, 2016 - 10:21 am
No absolute values of fit statistics can be compared using WLSMV. Chi-square difference testing is done using the DIFFTEST option.
Pia H.
posted on Tuesday, April 05, 2016 - 3:32 am
Dear Dr Muth�n
thank you for the quick reply. If have a follow-up question:
Is it possible to change the estimator or do I have to use the DIFFTEST manually? If the latter applies, can I have mPlus estimate the configural and the scalar model and then compare them using the DIFFTEST oder do I have to add the constraints in the manuscript manually and then use the DIFFTEST command?
Thank you very much,
Bengt O. Muthen
posted on Tuesday, April 05, 2016 - 6:35 am
Not sure what you are asking. You can't do DIFFTEST manually - too complex. You can estimate the model with ML and consider likelihood-ratio chi-square difference testing. But you won't get CFI with ML when you have categorical outcome which I think was your interest.
Pia H.
posted on Tuesday, April 05, 2016 - 6:44 am
Dear Dr Muth�n
I might have put too much emphasis on the CFIs in my first post. In fact, I only need to know if my model has configural and scalar measurement invariance.
Can I estimate two models using the model = configural command for the first model and model = scalar for the second one and then compare the two based on the chi square significance using the DIFFTEST command?
Thanks in advance,
Bengt O. Muthen
posted on Tuesday, April 05, 2016 - 6:46 am
Edwin Wouters
posted on Wednesday, July 27, 2016 - 8:34 am
Dear Sir or Madam,
I want to test the measurement equivalence of my factor (HIVOEXT) over two groups (1 = patients, 2 = others). Are these syntax correct to estimate the configural, metric and scalar invariance?
Thank you very much in advance
grouping is PatientCare (1 = pati 2 = others);
ESTIMATOR is MLR;
HIVOEXT by
HIVOEXT by
grouping is PatientCare (1 = pati 2 = others);
ESTIMATOR is MLR;
HIVOEXT by
!HIVOEXT by
grouping is PatientCare (1 = pati 2 = others);
ESTIMATOR is MLR;
HIVOEXT by
Linda K. Muthen
posted on Wednesday, July 27, 2016 - 11:49 am
The models to test for measurement invariance for continuous outcomes are shown in the Topic 1 course handout on the website. For categorical outcomes, see the Topic 2 course handout. Both are under the topic multiple group analysis. Chapter 14 describes the models for other situations.
Edwin Wouters
posted on Thursday, July 28, 2016 - 1:35 am
Dear Prof. Muthen,
Thank you for your quick reply.
I adjusted the syntax accordingly. Adding [HIVOEXT @0]; in the first part of the MODEL command for both configural and metric invariance, results in latent means fixed at zero for both groups. To test scalar invariance, I did not specify [HIVOEXT @0], causing the latent means to differ between groups.
This, however, feels counter-intuitive, as it seems I am freeing the means to obtain scalar invariance � whereas to my understanding the latent means should be fixed in this last model in order to obtain scalar invariance.
I am wondering whether I am mistaken. Should I only specify [HIVOEXT @0] in the scalar model and leave this specification out of the configural and metric model.
Thank very much in advance
Bengt O. Muthen
posted on Thursday, July 28, 2016 - 9:05 am
The factor means should be allowed to be different in the groups for the scalar model. It is still a much more restrictive model than metric and configural because you hold the intercepts equal across groups.
Edwin Wouters
posted on Wednesday, August 03, 2016 - 12:46 am
Dear Prof Muthen,
Thank you for your answer.
Consequently, should I allow the latent factor means to be different in the configure and metric model as well?
Thank you very much in advance
Bengt O. Muthen
posted on Wednesday, August 03, 2016 - 10:32 am
No, that would not be identified. Our UG shows how to set up these invariance models.
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text:
Well, you know it�s the Christmas season when the lawsuits start getting filed over nativity scenes on public property. Putting up these creches or manger scenes that depict the birth of Jesus in Bethlehem, has long been used during this holiday season to serve as a reminder of the reason behind all the newspaper inserts and television ads.
The first nativity scene is credited to St Francis of Assisi, in 1223, where he used real people and animals in a cave near Greccio, Italy. He set this scene to remind people that the meaning of Christmas was about Jesus and not materialism and gifts. Being approved by the Pope, other churches across Italy started setting up nativity scenes at Christmas. As the tradition spread, statues eventually replaced the humans and animals. But the meaning behind the nativity scenes was still to remind us of the reason Christmas. That is until the 1990, when nativity scenes started reminding us of the separation between church and state. Some people insisted that nativity scene should be in front of city halls or the town squares. This was not the reason St. Francis had when he established the first nativity scene. Nativity scenes are not the reason behind Christmas, they represent the meaning of Christmas and should be in front of churches, not court houses. It�s not the nativity scene that�s important, it�s the meaning behind it.
Some �Christians� would rather fight over a word rather than celebrate what the word stand for. Take, for example, the the phrase �Merry Christmas.� There are some who argue that if you say �Happy Holiday� your denigrating the meaning of Christmas. There is always talk of boycotting stores where �Happy Holiday� is used. Complaints are made over cities and states who erect Holiday Trees rather than Christmas Trees. What is overlooked is that the word �holiday� comes from the Old English words �Holy Day.� And, guess what, Christmas is a holy day. So saying Happy Holiday is saying happy holy day and the holy day is Christmas. It�s not the word that is significant, it�s the meaning behind it.
A few years back there was an uproar over the use of Xmas for Christmas. Some people claimed that this was �x�ing out Christ in Christmas. In fact, the church in the past had often referred to Christmas as Xmas because the Greek letter chi or �X� was the initial for Christ. You may have seen the Greek chi placed on the Greek letter rho, which looks like a capital �P.� This is a frequently used symbol for Christ on church vestments and hangings. It�s not the word, or letter, that�s important it�s the meaning behind it.
I suggest that rather than trying to put nativity scenes on public property, or avoiding places where they say Happy Holiday or have a Holiday tree, or complain about the �X� in Xmas, we start remembering the important thing that is behind all these things. Stop worrying about creches, trees, and terms and start worrying about the hungry, the poor, and the homeless. And have a Happy Holiday.

subset: commoncrawl


text:
Preschool camp offered
Published: Friday, June 21, 2013 at 10:50 p.m.
Last Modified: Friday, June 21, 2013 at 10:50 p.m.
Bricks4Kidz will host its first pre-school mini camp Thursday.
It runs from 6 p.m. to 7:15 p.m. Parents with children ages 3-5 are invited to accompany their children to learn about beach and ocean life, read stories, play and create a craft.
The cost is $80. Parents must attend. There are 15 spots available, and it costs $25 to hold a spot.
Other mini camps will be held July 11, 18 and 25. Bricks4Kidz is at 488 Corporate Drive in Houma. Call 872-2772.
Reader comments posted to this article may be published in our print edition. All rights reserved. This copyrighted material may not be re-published without permission. Links are encouraged.


text:
I love the work you lot have done, I was wondering as a cool feature would be to create web browser ablility similar to subsonic and weezo, but with the added feature that you can use mencoder or other codecs to stream and transcode either to a flash player or html5 player.
The reason being is that weezo and subsonic are only using ffmpeg and well MKV files with embedded subtitles do not show or other video files which have .srt files do not show either, (but weezo and subsonic can still view 10bit/Hi10P video).
It would be awesome if you guys could be able to create something like this with the added feature for user accounts and http internet streaming so, people can stream their music or videos when they are away from thier homes and let familys or friends use a guest account to view or stream music and video.
What do you think, plus I think, if you are able to do this you probably be the first http web media server with official subtitle ability.


text:
Illinois State Police District 2 -- See ISP Page
This page's information has been fully converted to the Wiki. Please click the WIKI Button for more information (Scanning Info, Websites, Unit IDs, Tones-Outs, Unidentified Frequencies, etc.)
*From a contributor - 3/15 - 460.325 D023, heard high power Emergency Dispatch near Elgin/South Elgin
WQLF265 - 151.0475 Repeater with 153.8975 input - Fire North - Burlington, Pingree Grove, Hampshire, Fox River [Patched to 154.385]
KNBG264 - 154.325 Repeater with 155.9175 input - Fire South - Big Rock, Kaneville, Maple Park [Patched to 154.07]
Repeaters on air around 10/9/13, patches should remain in effect until the end of the year, after which they will be used as backup/alternate fire channels.
*460.025, 460.075, 460.3875, add TRBO/NFM emissions for Judicial Center; 3/15 WPMH836 453.5375 Repeater (add DMR) Juvenile Justice System
*5/15 - also 155.37 Repeater and 155.6025 Repeater with inputs of 155.91 and 158.925.
*WQBJ776 - 158.835 Repeater with 159.24 input (possibly for county Department of Transportation)
*need coverage area and/or station location(s)
*WQQU270 - Maintenance/Security for 451.4375, 451.9625, 451.6375, 452.1375, 452.4875, 452.7875 (MOTOTRBO)
* 8/12 WNAA921 change to NXDN/NEXEDGE Trunked system, adding 461.6125 Repeater (at 1325 N Highland Ave)
Copely Hospital - Rt. 34 Aurora
*WQPP978 - Paging on 163.25
*WPQB423, changing 461.3375 R to 461.6875R, mobiles 464.95 466.025 466.05 466.075
*see the wiki for more info including radio prefixes
- Barrington Hills
- Carpentersville
- East Dundee
- Sleepy Hollow
- South Barrington
- West Dundee
**They are currently simulcasting dispatch for all of these agencies on at least three T-band frequencies.
155.7375 is the only frequency used by QuadComm in Kane County. QuadComm dispatches for Carpentersville Fire Department, East Dundee Fire Protection District, West Dundee Fire Department, Rutland Dundee Fire Protection District. No other agency specific frequencies are used today. All other communication is on the MABAS Fireground frequencies and/or IFERN, IREACH, etc.
Geneva Township Highway Dept (WQXE617): 151.160 - NBFM mobiles (new license 2/16)
Moecherville Water District NFP
Otter Creek Water District (S Elgin)
Rutland Township - KNEX220 - 151.13 Base/Mobile (NFM at Gilberts)
See McHenry County
*WQNI410 mobile extenders 770.0125 (800), 774.7375 (800). Probably will be used with the OpenSky system, Emission 12K19FW (not monitorable)
Aurora Public Library - WQUW660 - 451.6125 20-4wt NFM for Security/Ops
Dispatched by Montgomery Central Dispatch (Monrgomery Police Dept) on 154.0700 127.3 (same freq as Kane County fire south)
*WQPD462 - 151.01 BM (Montgomery), 159.135 BM (Aurora)
*WQPD462 - also 154.2125 Repeater with 151.1675 input (Montgomery-NFM)
As of 6 am 9/10/13, ATFD will be dispatched by KenCom Public Safety Dispatch in Kendall County.
ATFD will be switching to their repeater system and will no longer be using 154.07.
- 155.5425 (167.9) F1 Fire (repeats 153.8675)
- 154.1525 (162.2) F2 Fire (repeats 159.0675)
Unit numbers will be changing as follows:
*Chiefs - 611 is now 801, 612 is now 802, 613 is now 803, 614 is now 804
*Engines - 603 is now 821, 604 is now 823, 605 is now 824, 606 is now 822
*Medics - 651 is now 841, 652 is no longer in service, 653 is now 842
*Squad - 608 is now 871
Police dispatched by TRICOM
*Batavia Electric Department - 12/14 FCC Application 0006588043 for 453.2625 and 453.3375 Repeaters (DMR)
*need coverage area and station location
*Fire/Police dispatched by QUADCOM
*WQTZ987 - 154.025 Repeater with 158.82 input, 158.9925 mobiles (NFM)
*KRL411 Dundee Township - also 159.18 Repeater with 154.19 input (NFM/TRBO)
Police dispatched by QuadCom
Serves the Villages of East Dundee, South Barrington, Barrington Hills and unincorporated Kane & Cook Counties.
KSF846 - add 151.0325, 155.7375 Base/Mobile (NFM)
Police and Fire dispatched by TriComm
*WPHX763 - also 859.2875 Mobiles (WFM) possibly for additional fireground operations
*See the WIKI for additional information
*Police dispatched on/by Kane County Sheriff
Police/Fire dispatched by TRICOM
Hampshire Police - Dispatched by Kane County Sheriff
Hampshire Fire Protection District - Dispatched on Kane County Fire - North
Village located mostly in McHenry County
Huntley Fire Protection District
Village with a small portion located in western De Kalb County.
Police dispatched by/on Kane County Sheriff. Fire dispatched on Kane County Fire - North
*possibly part of Aurora Township FPD
*WQNW435 - EMA - 151.43 Repeater with 153.7625 input (NFM/TRBO)
Fire dispatched by Tri-Com on 151.4075 Repeater, which is simulcast with 154.7925 D174 Repeater. 159.99 is input to both. Repeater also shared with Sugar Grove.
Pingree Grove (& Countryside) Fire Protection District
Fire dispatched on 154.07 110.9 or Kane Fire North
*Dispatched on/by QuadCom 5
*Fire dispatched by TriCom
Police dispatched on/by QuadCom
*Police is now dispatched by KaneComm 159.150 (Kane County).
South Elgin-Countryside Fire Protection District - Dispatched by Elgin Dispatch (153.32)
South Elgin Parks and Recreation - WQJQ242 - Portables on 159.4575 (NFM)
*Police dispatched by Tri-Com 911 Central Dispatch - unit numbers "5xxx" on 470.9375
*Fire and Ambulance dispatched on Tri-Com Fire South Repeater
Sugar Grove FPD includes both the Village and Township of Sugar Grove
*WPLG344 - 461.5 Repeater for Maintenance/Security for Village Ops (Police Dept) (NFM) [Not in use, but license modified to narrowband in 2012]
Sugar Grove Fire Protection District
Police dispatched on/by Kane County Sheriff
Fire by West Chicago FPD and St. Charles Fire
*KCX409 add: 471.1625 R, 151.0325, 155.7375 BM, 159.0075 153.755 M
*Police/Fire dispatched by QuadCom
*Durham School Services - WQCP787 - 160.035 Base/Mobile (NFM at E Dundee)
Rosary High School - WQRS819 - 151.895, 151.955 15-5wt Portables; 151.715 is now Repeater wtih 158.385 input (NFM)
Cook Illinois Corporation - WIL448 - 472.9125 Repeater (at Aurora) NFM
*10/14 WQCS258 add mobiles on 451.5375, 456.3875, 456.4875, 456.5375, 456.58275 (DMR/NFM)
*also Prairie Knolls Middle School-Elgin, 461.5 Repeater; Burlington Central High School, 463.375 Repeater (DMR/NFM)
*see wiki for additional frequencies
*WQNB835 - 463.6125R (NFM); WQDY807 add 152.9975 Repeater with 158.3475 input (NFM)
Some locations located in Cook County
*see the wiki for additional frequencies
*WQNA401 - Security/Transportation - 460.2R (DMR/NFM at Maple Park)
*WQFG996 - 72.4 F GPS Based Clock System
*see the wiki for additional info and frequencies
*see the wiki for additional information
*see the wiki for more info


text:
A few months ago, I advanced THIS ARGUMENT:
It occurs to me that both sides in the debate over global warming have one thing in common: Hardly anybody really wants it to be true that climate-change trends pose a significant threat to our way of life.
But global-warming deniers want it to be untrue for reasons that are fundamentally different from mine, at least in one respect: They don�t just want to avoid the havoc of environmental catastrophe.� They find us exponents of mainstream scientific theories on climate change repugnant, and they want us to be disgraced. They have ...
Read more


text:
USA, 2012, 94 Minute Running Time
Genre/Subject: Comedy
Program: Special PresentationLanguage: English
Stand Up Guys stars Academy Award winners Al Pacino, Christopher Walken, and Alan Arkin in a tough but touching action comedy as retired gangsters who reunite for one epic last night.
Val (Pacino) is released from prison after serving 28 years for refusing to give up one of his close criminal associates. His best friend, Doc (Walken), is there to pick him up, and the two soon reteam with another old pal, Hirsch (Arkin). Their bond is as strong as ever, and the three reflect on freedom lost and gained, loyalties ebbed and flowed, and days of glory gone by. And despite their age, their capacity for mayhem is still very much alive and well—bullets fly as they make a hilariously valiant effort to compensate for the decades of crime, drugs, and sex they've missed.
But one of the friends is keeping a dangerous secret—he's been put in an impossible quandary by a former mob boss, and his time to find an acceptable alternative is running out. As the sun rises on the guys' legendary reunion, their position becomes more and more desperate and they finally confront their past once and for all.
Click here for Information on Group Rates and Discounts!
DIRECTOR: Fisher Stevens
Producer: Sidney Kimmel, Tom Rosenbeg, Gary Lucchesi, Jim TauberEditor: Mark LivolsiScreenwriter: Noah HaidleCinematographer: Michael GradyPrincipal Cast: Al Pacino, Christopher Walken, Alan Arkin, Julianna Margulies
CLICK IMAGE for the Member Newsletter
CLICK IMAGE to submit to the Denver Film Festival. Withoutabox logos are trademarks of Withoutabox, a DBA of IMDb.com Inc. or its affiliates.
CLICK IMAGE to Renew or JoinMembership Updates and Sign Up
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Site powered by: Luminate


text:
Chapters of the proposed 2012 update to the Eagle River Watershed Plan are now available for public review and comment at www.eaglecounty.us.
The 30-day review period is underway and will continue through Aug. 18. The executive summary and appendices will be finalized during the public hearing process, anticipated to take place in September.
All comments received will be provided to the Eagle County Planning Commission for consideration during public adoption hearings. Comments can be submitted by mail to Cliff Simonton, c/o Eagle County Community Development, P.O. Box 179, Eagle, CO 81631; or by phone to (970) 328-8751; or by email to firstname.lastname@example.com.
First adopted in 1996, the Eagle River Watershed Plan provides information on the status of streams and rivers in the county, and outlines a collaborative philosophy for protecting and improving water quantity, water quality, wildlife habitat and water-related recreational opportunities. Work to update the plan began in 2010, in recognition of the changes that occurred in the county during the previous 16 years.


text:
ACE Certificate
In today's competitive marketplace, employers and graduate schools require more than simply filling out online applications and viewing websites.
At UE, we offer every student the tools to give you more.
The Advanced Career Education (ACE) Certificate program offers the opportunity to gain the skills needed to easily communicate your professionalism, skills, and career focus to potential employers and graduate schools. With a partnership that includes faculty, advisors, the Center for Career Development, experiential education opportunities, and an extensive network of alumni and friends, the ACE Certificate program will challenge you to evaluate your career goals, polish your professional image, and create the building blocks of a lifelong professional network.
The program also meets the requirements of the INdorsed Career Ready Graduate Program, initiated by the Career Development Professionals of Indiana, an association dedicated to promoting the highest possible outcomes for college graduates in Indiana.
For more information or to register, visit the ACE Certificate page of our online Career Center.


text:
2011 Annual Report
5)% RH and 16:8 (D:N) light photoperiod until they started active feeding (usually about two to three weeks after field collection). Only laboratory conditioned adults were used for the insecticide bioassays. Adult BMSB were separated based on gender and grouped by five individuals. Each group was placed in a plastic 0.473 l (16.0 oz) clear plastic container supplied with a small piece of carrot (2 mm slice) as a source of food and moisture. Adult BMSB in six containers per gender (n=30 males or females) were separated and treated with a topical application of water solution of formulated insecticide. Each individual BMSB adult received 2 ul drop on the upper dorsal (just below pronotum, on the edge of scutellum) surface of the exoskeleton. The concentration of each insecticide solution was determined as equivalent of the recommended field rate of the product assuming application of 100 gal (378 l) of spray solution per acre (0.4 ha). A biological surfactant, Triton-X 100, was also added to each solution (1 ml of surfactant per 100 ml of solution). Water only plus surfactant was used as a control treatment. Treated BMSB individuals were kept in enclosed containers at 23 C and 16:8 photoperiod. The mortality of BMSB was assessed at 4, 24, 48, 72, 96 and 120 hours after the application. During each evaluation, each individual insect was scored as either live, dead, or moribund. The moribund category includes individuals unable to move forward or backward in a coordinated manner but still able to respond to touch by a camel brush by moving legs or antennae. Close to 60 individual combination of various products/rates were tested up to date with some products being tested multiple times at various field rates or available formulations. The results of the bioassays were made available to stakeholders throughout a series of meetings and extension publications presented during the spring and summer of 2011. The ADODR has monitored activities through emails, site visits, and calls.


text:
The question is Why was there a witch craze in the C17th and for my target grade it says: Your work was analytical & detailed, listing and explaining in some detail both general & specific reasons why people were accused of being witches, ascribing significance &basing conclusions on evidence. I already have some reasons for why people were accused of being witches like religion and the fact that people couldn't explain things by science but i don't really know what to do and how to do it. we also have to enclude some source material but i don't know where to get any. Again I would be very grateful if you could help me. I would ask my teacher but it is ½ term. Thank you again!
Edited by lollypop, 19 February 2012 - 11:36 AM.


text:
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Very interesting article,This is the first time I hear about 2 different types of ICCA. Still reading and just in time for the medical conference for tomorrow.
Thanks and
God bless.
I also didn't know about different classes of ICC. Thanks for the study article.
this may explain the complexity of ICCA subtypes.
Intrahepatic Cholangiocarcinoma
An increasing global incidence of ICC [5] has recently hastened research in this field to understand the mechanisms underlying pathogenesis of this dreadful disease. Reviewing the mechanisms of ICC indicates that similarities can be drawn between ICC and HCC, which may improve the prospects of this disease in a clinical setting. Particularly, the tumorigenesis models proposed for ICC development are remarkably similar to those for HCC. Furthermore, several histopathologic and gene expression profiling studies have shown PLC tumors that exhibit a combination of HCC and CCA traits, suggesting an overlap between these tumor types. A subtype of tumors showing combined characteristics of hepatocellular-cholangiocarcinoma (CHC) have been reported and proposed to develop from the bi-potential liver stem cells [97]. Even more recently, a new subtype - cholangiocarcinoma-like HCC (CLHCC) -was discovered and characterized as HCC expressing CCA-like traits. The heterogeneity observed between all 4 tumor subtypes could be indicative of their cellular origins from different developmental stages and may also represent a novel way to approach targeted therapy in CCA and HCC [98]. from the link below.
God bless.
Thanks for that Linda. I'll go put that link over on the FB page as well.
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- Nov 15:
- NCS Division V: Valley Christian 50, St. Bernard Catholic-Eureka 18
- NCS Division III: El Cerrito 64, Tamalpais 8
- NCS Division I: James Logan 14, Liberty 7
- NCS Division I: Granada 24, Amador Valley 20
- NCS Division II: Montgomery-Santa Rosa 27, American 0
- NCS Division I: California 37, Dougherty Valley 12
- NCS Division II: Clayton Valley Charter 62, San Lorenzo 7
- Nov 15:
- Deer Valley romps over Castro Valley in playoff opener
- East Bay high school/community college scoreboard for Friday, Nov. 15 events
- NCS Division I: Pittsburg 47, Irvington 13
- Friday's NCS high school football scores
WALNUT CREEK -- Once the Northgate High football team's defense finally produced a stop against College Park, Friday's North Coast Section Division II playoff opener turned in the Broncos' favor.
Down by nine points in the second half, Northgate needed a fourth quarter comeback to beat College Park 41-29 and earn a quarterfinal date with the winner of Saturday's game between Miramonte and Tennyson.
The No. 6 seeded Broncos trailed at halftime and after three quarters, but scored the final 21 points of the game when its defense finally started clicking.
Offensively, Northgate (9-2) relied on a heavy dose of Eric Haynes, who rushed for 263 yards and four touchdowns on 27 carries. His final score of the night, a 5-yard run in which he spun off defenders into the end zone, put the Broncos ahead for good with 9:31 left.
For three and a half quarters, Northgate's defense had no answers for No. 11 College Park, which led by as much as 29-20.
But when the Falcons (6-4-1) gambled on a fourth-and-one from their own 26 yard-line, the move backfired. Trevor Sprung was stuffed at the line of scrimmage, and the Broncos scored three plays later when Kyle Austin connected with Nick Mastrelli. The touchdown pulled the Broncos to within 29-27.
Northgate went on to score on their next two possessions to complete the comeback win.
Haynes scored Northgate first three touchdowns of the night, tallying 172 of his yards in the first half.
"Guys key on him and know he's going to run it," said Northgate coach Justin Lowell. "He's a physical guy and he's able to take that beating and keep going game in and game out. He did a great job tonight, especially when we kind of sputtered offensively, he was always able to get us a first down when we needed it."
Northgate had defeated College Park 49-14 in Diablo Valley Athletic League play less than a month earlier, but the rematch played out much differently.
"They were more prepared for us," Haynes said. "They knew what we like to do and came out with a game plan to stop us."
College Park headed into the locker room up 22-20 after a back and forth first half. Timmy Jackson threw a 43-yard touchdown to Michael Sabeti on a halfback option pass, and sophomore David Jones returned a kickoff 95 yards for a score in the wild opening half.
The Falcons' defense also produced goal-line stand, stopping Haynes half a yard away from the end zone just before the break.